an assessment of the sustainability of current account imbalances in

AN ASSESSMENT OF THE SUSTAINABILITY
OF CURRENT ACCOUNT IMBALANCES
IN OECD COUNTRIES
Autores: Mariam Camarero(1)(*)
Josep Lluís Carrion-i-Silvestre(2)
Cecilio Tamarit(3)
P. T. N.o 16/09
(1) Department of Economics. Jaume I University.
(2) Department of Econometrics, Statistics and Spanish Economy. University of Barcelona.
(3) Department of Applied Economics II. University of Valencia.
(*) Corresponding author: Department of Economics, Jaume I University, Campus Riu Sec, E­
12071 Castellon (Spain). Phone: 34+964728595. Fax: 34+964728591. e-mail:
camarero@eco.uji.es; http://www3.uji.es/~camarero
N.B.: Las opiniones expresadas en este trabajo son de la exclusiva responsabilidad de los
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Edita: Instituto de Estudios Fiscales N.I.P.O.: 602-09-006-9 I.S.S.N.: 1578-0252 Depósito Legal: M-23772-2001
INDEX 1. INTRODUCTION
2. SOME STYLIZED FACTS ABOUT TH CURRENT ACCOUNT
3. BRIEF EMPIRICAL LITERATURE REVIEW
4.
1.
1.
1.
1.
1.
1.
THEORETICAL FRAMEWORK
4.1. The classical flow equilibrium approach: sustainability of the current
1.1. account and the intertemporal budget constraint
4.2. The stock approach: the arithmetic of intertemporal solvency (Net
1.1. international debt to GDP ratio)
4.3. The unified approach of Gourinchas and Rey (2007): foreign debt and
1.1. the current account
5. ECONOMETRIC METHODOLOGY AND RESULTS
1.
1.
1.
1.
1.
1.
1.
5.1.
5.1.
1.1.
1.1.
1.1.
1.1.
1.1.
Testing for current account sustainability and external debt solvency:
panel analysis
5.1.1. Testing for the presence of multiple structural breaks
5.1.2. Testing 1(0) stationary on individual time series
5.1.3. The issue of cross-section independence
5.1.4. Panel data tests with cross-section dependence and structural
5.1.4. breaks
1. 5.2. Testing for the unified approach
1. 1.1. 5.2.1. Bai-Perron estimation results
1. 1.1. 5.2.2. VAR and impulse-response results: the case of US and Spain
6. CONCLUSIONS
REFERENCES
APPENDIX
SÍNTESIS. Principales implicaciones de política económica
—3—
ABSTRACT
In this paper we analyze the solvency and the sustainability of the current
account for a group of twenty OECD countries using panel data methods. We
test for the main hypotheses that have been formulated following both the flow
and stock approaches in an intertemporal setting. Moreover, we also formulate
a unified testable model based on Gourinchas and Rey(2007). The main results
of the models can be tested using stationarity tests. For this purpose, we apply
panel stationarity tests allowing for structural breaks and cross-section
dependence. The evidence points to the solvency of the external accounts;
however, the two variables evolve around a shifting deterministic component
implying, hence, the non sustainability of the external position in most of the
countries considered. We estimate, also allowing for structural breaks, the
reduced-form parameters linking the two variables, following Gourinchas and
Rey. This relationships is estimated for the sub-periods defined by the breaks.
For the majority of the countries and time-periods the parameters are positive,
smaller than one and significant, as expected. Concerning the dynamics, we find
that solvency is reconvered after major shocks affecting the countries' external
accounts.
Keywords: Current account, panel data, structural breaks, cross-section
dependence.
JEL codes: F32, F41, C23.
—5—
Instituto de Estudios Fiscales
1. INTRODUCTION
Since the beginning of the 1990s, current account (CA) imbalances have been
widening considerably in the world economy. Economic globalization has meant
an increase in international trade and capital mobility facilitating the financing of
larger and more persistent current account imbalances. Among the OECD
countries there is a clear trend toward larger imbalances, i.e. by 2007, the
current account imbalances, whether surplus or deficit, of the OECD countries
were more than twice as large as in 1988. However, the trend towards large
imbalances is not confined to the OECD countries. These imbalances have been
more acute between China and the oil exporter countries, on the one hand, and
the US, on the other. Many emerging economies now show larger surpluses in
their current accounts, although it may be necessary to distinguish between
those who are enjoying a temporary surplus due to a favorable movement in the
prices of their exports (as in the recent run-up in commodity prices) and those
whose surpluses are the result of the pursuit of a particular development
strategy. According to the World Trade Report 2008 (WTO, 2008), emerging
East Asia has followed an export-led development strategy which was
supported by exchange rate policies that anchored domestic currencies to the
US dollar. It has been a successful development strategy resulting in the rapid
mobilization and employment of tens of millions of workers. The means to bring
this about is the cross-border transfer of goods and services to the centre
country in exchange for financing its deficits (Dooley et al., 2007).
The flow of savings to developed countries has also been encouraged by the
lack of financial and capital market development in emerging Asian economies.
The underdeveloped nature of the domestic financial or capital markets has
become a bottleneck preventing the effective channeling of domestic savings
into worthwhile investment projects at home. But the size of the imbalances has
raised the key question of their sustainability and the nature of the adjustment
process. Therefore, there has been a renewed interest in the study of the
determinants of the dynamic adjustment of external imbalances. In part, larger
current account imbalances reflect the impact of greater capital and financial
market integration. A current account deficit reflects dissaving by domestic
residents, an excess of absorption over income. The fact that it is occurring
reflects a willingness by foreigners to finance that excess absorption by
accumulating future claims on the earnings of domestic residents. As a
consequence, net foreign liabilities have also been growing, generating concern
that policy measures may be required if costly and destabilizing shifts in market
sentiment are to be avoided.
—7—
The weight of experts' opinion suggests that these imbalances will ultimately
decline although there is no consensus on when or on the manner, whether
smoothly or abruptly, in which it would occur (Clarida, 2007). But there seems
to be broad agreement that some combination of exchange rate and asset price
changes would play a role during the process of adjustment. Studies of past
adjustments in industrial countries point to the challenges ahead. Larger deficits
take longer to adjust and are associated with significantly slower income growth
during the current account recovery (Freund and Warnock, 2007).
Consumption-driven current account deficits involve significantly larger
depreciations than deficits financing investment. Obstfeld and Rogoff (2006)
suggest that a large depreciation of the US dollar, something in the order of 30
per cent, could accompany the process.
While temporary current account deficits may simply reflect the reallocation
of capital to countries where capital is more productive, persistent deficits may
be regarded as more serious. Deficits may lead to increased domestic interest
rates to attract foreign capital. However, the accumulation of external debt due
to persistent deficits will imply increasing interest payments that impose an
excess burden on future generations. Now, adjustments to large current
account imbalances are complex processes. The speed and economic effects
depend on many factors. How much of the adjustment takes place through
changes in asset valuation? How much through a reduction in absorption? How
much in the form of expenditure switching? It will also matter how much
international coordination among financial and central bank authorities takes
place to ensure a supportive policy environment. Thus, the discussion above
should not be seen as simplifying the challenges that are involved. If one can take
a specific example, the soft-landing scenario requires that the acceleration of US
export growth be matched by increased demand for US goods from the rest of
the world. This would need to be triggered by just the right kinds of movements
in exchange rates, asset and goods prices.
Mann (2002) considers that sustainability should be viewed both from the
domestic and international finance point of view. A sustainable current account
is one that does not trigger feedback effects on domestic variables (investment
and savings) or does not lead to significant international portfolio reallocations
leading to changes in interest rates. We can distinguish three approaches in the
theoretical literature that analyzes the current account balance. First, the
conventional non-optimizing models, that comprises the Keynesian and
monetary views, generally using reduced-form solutions and examining
aggregated macroeconomic aspects. Although these models à la Mundell­
Fleming-Dornbusch provide a useful policy framework, the main drawback is that
they are not based on microeconomic foundations and optimizing behavior of
the economic agents. A second approach is the micro-founded intertemporal
optimizing models developed in the 1980's that use the intertemporal budget
—8—
Instituto de Estudios Fiscales
constraint. The major advantage of these models is that they deal with current
and capital account behavior simultaneously through direct and portfolio
investment flows across border along with trade in goods and services. The use
of these models has facilitated the analysis of the sustainability of current
account deficits. The intertemporal models developed until the late 1980's
generally assumed perfectly flexible domestic prices and ignored the short-term
price rigidities in product and factor markets. Finally, a third theoretical
approach is the extension of the intertemporal models developed during the 1990's
that introduced nominal rigidities and market imperfections into the dynamic
general equilibrium models, being the Obstfeld-Rogoff Redux model the major
milestone in the intertemporal approach to open-economy macroeconomics.
These models provide a sound micro-theoretical framework, although they
lack a matching empirical validation of the theoretical propositions. The
empirical content in some of the models remains restricted to only calibrated
simulations. The policy formulations at the central banks, government
organizations, International Monetary Fund and the World Bank require an
empirically tractable and econometrically estimable model to verify the
theoretical propositions.
More recently, some studies have extended the modern portfolio
optimization theory to the current account and suggest that the marginal unit of
wealth arising from a positive productivity shock is allocated according to the
existing portfolio choices, and that changes in saving lead to changes in current
account proportional to the share of foreign assets in total assets.
Kraay and Ventura (2003) suggest that, in the long run, countries invest a
marginal unit of saving in domestic and foreign assets in the same proportions as
in their initial portfolios. In the short run, countries invest a marginal unit of
saving mostly in foreign assets, and only gradually do they rebalance their
portfolio back to its original composition. Countries not only try to smooth
consumption, but also domestic investment, and they use foreign assets as a
buffer stock.
Lane and Milesi-Ferretti (2001, 2002) have examined the relationship
between current account and changes in net foreign asset position at market
value, and showed that the correlation between them is low or even negative.
Lane and Milesi-Ferretti (2004) suggest that currency fluctuations influence the
rates of return on inherited stocks of foreign assets and liabilities, in addition to
operating through the traditional trade adjustment channel. The large gross
cross-holdings of foreign assets and liabilities suggest that the valuation channel
of exchange rate adjustment has grown in importance, relative to the traditional
trade balance channel. More recently, Gourinchas and Rey (2007) have
decomposed the external adjustment into a financial (valuation) channel and a
trade (net export) channel and show that the deteriorations in net exports or
—9—
net foreign asset position of a country have to be matched either by future net
export growth (trade adjustment channel) or by future increases in the returns
of net foreign asset portfolio (financial adjustment channel). The valuation
channel is important in the medium-term and the net export channel is
important in a long-time horizon.
The aim of this research is to test for sustainability following the framework
defined in Milessi-Ferretti and Razin (1996) and Taylor (2002). According to this
stream of the literature, it is possible to define two key concepts regarding the
stochastic properties of the current account. First, the current account is said to
be solvent if it is I(0) stationary. Second, the current account is sustainable if the
economy is able to satisfy its long-run intertemporal budget constraint without a
drastic change in private sector behavior or policy shifts. This is a more general
concept and does not depend on any particular model. At the same time this
concept of sustainability is a sufficient condition for other concepts to hold, with
the advantage of its easy testability. According to Trehan and Walsh (1991),
current account stationarity is a sufficient condition for the intertemporal budget
constraint to hold. To this aim we first test for stationarity of two variables: the
Current Account (CA) to GDP ratio and the Net Foreign Assets (NFA) position
to GDP ratio. The first variable is representative of the traditional flow approach
to the intertemporal budget constraint, while the second, provided that we use
the stocks build by Lane and Milesi-Ferretti (2007), that consider the valuation
effects in the financial markets, is already a methodological improvement
compared to previous empirical work. Finally, as a second step we analyze a key
relationship for the long run stability of the unified approach model of the
current account adjustment as defined in Gourinchas and Rey (2007).
For this purpose we use a panel data unit root test that allows for the
presence of structural breaks and cross-section dependence. From an
econometric point of view the contribution of this paper is twofold. First, we
test for the presence of structural breaks affecting the CA time series,
considering as a particular case the situation with no structural breaks. Once the
presence of structural breaks has been investigated, then individual stationarity
test statistics are computed. Second, such individual tests can be pooled to
define panel data based test statistics, which permit an assessment of the CA
stochastic properties using more powerful statistical tools. The statistical
inference is conducted taking into account the presence of cross-section
dependence through the computation of the bootstrap distribution and the use
of approximate common factor models. As for the third variable analyzed, that
is, the relationships between CA and NFA we first, use the Bai and Perron
(1988) methodology in order to ascertain the possible structural breaks in the
relationship on a country by country basis. It is worth mentioning that the
application of the Bai-Perron methodology to estimate the number and position
of the structural breaks requires the variables under analysis to be stationary in
— 10 — Instituto de Estudios Fiscales
variance, which is consistent with the null hypothesis and previous findings of
the univariate analysis of the two ratios. Secondly, we analyze the relationship
between both variables accounting for possible co-breaking relations in a
dynamic heterogeneous panel setting.
The remainder of the paper is organized as follows. Section 2 develops the
present global imbalances situation in the economic relations at the world level,
describing the main stylized facts. Section 3 displays a revision of the previous
empirical literature, emphasizing the main issues related to the relationship
between increasing economic integration and the external imbalances. In
Section 4 we discuss the theoretical framework that guides our empirical
investigation on the mechanisms of international financial adjustment. Section 5
presents our econometric methodology and describes the construction of our
annual database for the OECD countries. The empirical results are presented in
Section 6 and, finally, Section 7 concludes.
2. SOME STYLIZED FACTS ABOUT THE CURRENT ACCOUNT
Prior to the empirical analysis developed in the next sections we will study
the stylized facts associated with the current account and the external position
of the developed countries.
The first stylized fact that emerges is the intense degree of financial
globalization that has occurred during the last decade, despite the financial crises
and the reversal in global stock markets values in 2001-2002. During the last
decade, the amount of financial wealth has increased steady in the world. The
intensity of the process can be understood if we compare Figures 1 and Figures
2 and 3. Figure 1 shows the sum of exports and imports as a percentage of GDP
for the Euro area countries, the UK, the US and Japan. Trade openness has
increased steadily during the period considered. Even if the US and Japan are
less open than the UK or the EMU countries, all of them have doubled,
approximately, their level. This is in sharp contrast with Figures 2 and 3, where
financial integration is measured as the sum of foreign assets and liabilities as a
percentage of GDP, as proposed by Lane and Milesi-Ferretti (2008). We have
presented this index selecting the countries according to criteria linked to their
size and characteristics. The Southern and peripheral European countries (Spain,
Ireland, Italy, Greece, and Portugal) are shown in the left graph of Figure 2.
With the exception of Ireland (an outlier in comparison with the rest, due to its
exceptionally high level), these countries have increased their financial
integration around nine or ten times. The evolution is different in the graph on
the right: larger Anglo-Saxon countries, such as the US, Canada, Australia and
New Zealand, are four or five times more financially integrated than at the
— 11 — beginning of the seventies. The exception is the UK in this case, a major financial
center, especially after the advent of the euro. The behavior of the countries
depicted in Figure 3 turns out to be similar to the other European countries
mentioned above: financial integration has augmented around ten times.
Thus, even if international trade has increased substantially, on the financial
side international capital flows have expanded even more rapidly, and the
financial linkages, as well as the real ones, have tightened across countries. A
second stylized fact is that this process has caused important external
imbalances in the current account. Figures 4 and 5 show the current account
balance as a percentage of GDP for the same country groups as above1. In
Figure 4, on the left graph, the peripheral Southern European countries have
experienced significant deficits. The most dangerous positions, with deficits
above 10% of GDP, are those of Portugal and Spain (and to a lesser extent,
Greece); however, Italy and Ireland also worsened their balance after 2000.
According to Blanchard (2007), these very large deficits in rich countries reflect
mostly private saving and investment decisions. The question is why these
countries have been able to experience such deficits without having suffered a
reversal and, therefore, an adjustment. A possible explanation is that in a
monetary union, the broad external balance of the European economy hides
significant differences in external positions across individual European countries.
Some of them, such as the Southern peripheral European countries are
converging towards the core EMU countries. The external constraint that
individual countries may face is not longer working in a monetary union.
However, according to Lane and Milesi-Ferretti (2007), the exposures across
Europe are very heterogeneous (differences in trade patterns, financial
exposures, and net external positions) so that the process of adjustment may
constitute an asymmetric shock. This implies bilateral real exchange rate
adjustments between creditor and debtor countries as members of the Euro
area. This heterogeneity can be observed in Figure 5: with the exception of
France, after the euro, the “core” euro-area countries and the Nordic economies
maintained a surplus in their current account balance. Therefore, some members
of the EMU may experience significant and probably painful adjustments in their
external position when the credit conditions become tighter.
However, it should be emphasized that the EMU countries are not the only
ones to have been affected, in recent years, by external imbalances. The case of
the US has been discussed abundantly due to its magnitude and the persistence
of the creditor position. Other OECD countries, such as New Zealand and
Australia are also experiencing similar imbalances.
1
Caballero et al. (2008) divide the world into four groups: the United States (and similar
economies such as Australia and the United Kingdom); the Euro Zone; Japan; and the rest of
the world. This classification also emerges from our stylized facts analysis.
— 12 — Instituto de Estudios Fiscales
An alternative, although complementary, approach to the nature and
dimension of external imbalances can be gathered by looking at the net foreign
assets (NFA hereafter) position of this same group of countries. Figures 6 and 7
show quite clearly another stylized fact: the preeminence or the persistence of
the net debtor positions among the developed countries. The only exceptions
are Japan, Norway (the only oil exporting country in the sample), and a group of
core EMU members (Germany, France and Belgium). The negative values of the
NFA position (sometimes reaching 50% of GDP or even 75% as in Spain)
reflect the cumulated effect of persistent current account deficits and therefore,
the imbalance between foreign assets and liabilities. Many rich countries have
benefited from the high degree of international financial globalization and have
been able to finance their growing current account imbalances through foreign
capital entries. However, the deterioration of the NFA position has been severe
in many cases and calls for painful adjustments.
3. BRIEF EMPIRICAL LITERATURE REVIEW
As the current account represents the rate at which a country accumulates
or decumulates foreign assets, one approach to judging whether an external
balance of a given size is a problem or not is to see whether it is consistent with
the assumption that all external debts will ultimately be repaid. This is the
notion of intertemporal solvency. This concept, however, is a relatively weak
criterion as far as giving warning of an emerging problem. The reason is that
solvency requires only that, in the very long run, all debts be repaid. Since this is
equivalent to saying that large trade deficits today will be offset by equally (in
present value terms) large trade surpluses in some future period, a country can
remain technically solvent even while running large external deficits as long as
policies are adjusted as needed in the future to bring about the required
surpluses that enable debts to be repaid. Therefore, it can be argued that
intertemporal solvency imposes too few restrictions on the evolution of the
current account and external debt over the medium term to be of much
operational value in telling us when a country's external position warrants
attention from policy makers.
A more demanding criterion is sustainability. This concept adds on to the
notion of solvency the idea that policies remain constant for the indefinite
future. Thus, an external position is sustainable if, under the assumption that
policies do not change, the country does not violate its intertemporal solvency
constraint. The problem with the sustainability concept is that what matters
for the current account are people's expectations of future policies rather
than the policies themselves. These expectations are notoriously difficult to
— 13 — observe and measure, which makes the sustainability concept difficult to apply
operationally.
Economists do not agree on a precise definition of a sustainable current
account. In general, sustainability refers to a stable state in which a current
account deficit generates no economic forces of its own to change its trajectory.
In this research, a country's current account deficit is defined as unsustainable
when it triggers a sharp hike in domestic interest rates, a rapid depreciation or
some other abrupt domestic or global economic disruption. Using this
definition, a sustainable current account is one that changes in an orderly fashion
through market forces without causing jarring movements in other economic
variables, such as the exchange rate.
The traditional Keynesian approach to the current account put the emphasis
on international price competitiveness and relative demand in explaining current
account movements. However, the intertemporal approach that appeared from
the beginning of the 1980's has emphasized the role of forward-looking
expectations in explaining current account patterns. The current account of a
country is treated as a reflection of consumption and investment decisions that
span over long-term horizons. Thus, the standard intertemporal model of the
current account considers the current account from the saving-investment
perspective and features an infinitely lived representative agent who smooths
consumption over time by lending or borrowing abroad. As the global
integration of the financial markets increased from mid 70's, there was a rapid
expansion of two-way capital flows and gross external asset and liability
positions that contributed to the creation and sustainability of current account
imbalances. Therefore, the intertemporal approach became a more appropriate
framework to analyze the dynamics of the current account.
The intertemporal approach to the current account stresses that, since the
current account is the difference between national saving and investment,
external deficits or surpluses result from intertemporal investment and
consumption decisions by firms, households and the government2. Thus, when
international markets provide limited insurance opportunities, borrowing and
lending enables economic agents to smooth consumption through intertemporal
trade, enhancing economic efficiency. The empirical applications of this
approach evolved along two main lines of research.
The first strand of the literature applied the present value test, as developed
by Campbell and Shiller (1987). Under some simplifying assumptions and using a
methodology developed by these authors in a different context, one can
estimate the current account series that would have been optimal from a
consumption smoothing perspective. The standard model implication is that the
2
See Obstfeld and Rogoff (1995, 1996) for a survey of the literature.
— 14 — Instituto de Estudios Fiscales
current account balance equals the present value of expected future declines in
net output (output less investment and government spending). The intertemporal
approach to the current account was first popularized by Sachs (1981) and
considers net accumulation of foreign assets as a way for domestic residents to
smooth consumption intertemporally in the face of idiosyncratic income shocks.
Namely, in response to positive temporary shocks to net output, domestic
households can increase both current and future consumption by lending
internationally, either directly or through financial institutions. Conversely, in
response to permanent shocks that raise net output in the long-run by more than
in the short run, domestic households can optimally smooth consumption by
borrowing in the international financial markets. To the extent that the
permanent increase in net output is driven by shocks to productivity, borrowing
in international financial markets allow the domestic economy to sustain higher
rates of domestic investment without cutting current consumption.
For more that two decades, these basic propositions have been tested using
variants of the present-value model originally conceived by Campbell (1987) and
Campbell and Shiller (1987) with mixed results. Starting with Ahmed (1986) and
Sheffrin and Woo (1990), economists have compared actual current account
data with this optimal benchmark leading to the general result that while the
model-predicted current account is positively correlated with the actual series,
the latter is substantially more volatile, what implies a statistical rejection of the
model. Although the positive correlation means that consumption-smoothing
plays a role in the dynamics of the current account, the finding of excess current
account volatility has been used to reject the proposition of limited international
capital mobility, as stated by Feldstein and Horioka. The present value
framework was then extended in several directions in more recent papers.
These studies have tried to generate extra predicted volatility through real
exchange rates and interest rates variability (Bergin and Sheffrin, 2000), by
incorporating consumption habits (Gruber, 2004) or by adding an exogenous
world real interest rate shock (Nason and Rogers, 2006). The extent to which
the model performance is driven by the empirical failure of the auxiliary
assumptions commonly adopted to make the model testable is unclear but has
been claimed as the main reason for that. In addition, present-value tests do not
distinguish between temporary or permanent shocks driving the dynamics of a
country's net foreign liabilities.
The second strand of the literature has applied standard econometric
techniques to establish if there is a long-term relationship between the current
account and macroeconomic fundamentals – i.e. relative GDP per capita, the
demographic structure or fiscal policy3. Recent literature addressing these issues
3
See, for example, Debelle and Faruquee (1996), Chinn and Prasad (2003) or Bussiere et al.
(2004).
— 15 — has used DSGE models with non conclusive results4. Moreover, due to the lack
of a precise definition, no universally accepted measure of sustainability exists.
Many economists gauge sustainability by examining the value of a country's
external obligations. In this context, two commonly used measures are the ratio
of the country's current account deficit to GDP and the ratio of the country's
net international debt to GDP. Insight into the causes of the deficit can be
gained by looking at how the deficit is financed. Examining the ratio of net
international debt to GDP provides an alternative method for assessing the
sustainability of a country's current account deficit. Net international debt is the
accumulation over time of current account deficits. If an economy runs a
current account deficit consistently, net international debt may become so great
that foreign investors lose confidence in the economy's ability to service its debt
or, worse yet, repay the principal. Once this happens, interests rates must rise
or the borrowing country's currency must depreciate to enable the country to
continue financing its deficit. In this case, the current account deficit has
generated economic forces of its own to change its trajectory, and the current
account deficit and the associated debt have become unsustainable. In balance
of payments terminology, net capital inflow is the financial counterpart of the
current account deficit. Thus, current account positions which appear justified
from such perspective can only materialize subject to the constraints implied by
international capital flows. In other words, a country that is solvent may
nevertheless not be able to finance a particular current account deficit if investors
are not willing to provide the required funds, i.e. if the country is liquidity
constrained. However, recent empirical literature trying to test for this approach
still relies only on flows to assess the dynamics of the adjustment process5.
From a theoretical perspective, the above flow approaches have a major
drawback, as they ignore valuation effects of stocks of foreign assets and
liabilities and assume that the current level of net foreign assets (NFA) is
sustainable. Although this mechanism could help to a gradual rebalancing, these
benefits could turn into a problem if policies are not consistent with a credible
medium-term policy framework aimed at external and internal balances, as
expectations may not be well anchored. In this case, investor preferences may
quickly change and the fallout from disruptive financial market turbulence would
likely be more elevated than it had been otherwise. Moreover, a country
4
See, for instance, Blanchard and Giavazzi (2002), Fagan and Gaspar (2007) or Bems and
Schellekens (2007).
5
For example, Bussière et al. (2004) extend the standard intertemporal model by introducing
habit formation and non-ricardian consumers to account for current account behavior in the
OECD and in EU acceding countries. Similarly, Zanghieri (2004) extends this analysis by
projecting the future level of debt using the forecasts of current account minus FDI flows.
Depending on the assumed share of FDI in the current account deficit, CEECs' debt will be
stabilized (high share of FDI) or will continue to grow (low share of FDI).
— 16 — Instituto de Estudios Fiscales
running persistent current account deficits might be at the same time improving
its NFA position if capital gains on its foreign assets exceed those on its foreign
liabilities (Lane and Milesi-Ferretti, 2006). Additionally, if the country is located
away from its equilibrium level of NFA, the current account deficit can be
sustained precisely because the economy is adjusting to a higher level of long­
term liabilities. Edwards (2001) shows that this adjustment process can lead to
quite substantial current account deficits.
In our view, a stock approach can cope successfully with this problem. Stocks
are also less volatile and can provide long term relationships that are easier to
estimate. The stock approach has recently been used by several authors thanks
to the development of an external wealth database by Lane and Milesi-Ferretti
(2006). They use their own estimates of external positions to study the
determinants of NFA in developing and industrial countries and they find that
public debt, GDP per capita and a set of demographic variables give a good
account of the patterns of external holdings (Lane and Milesi-Ferretti, 2001).
Calderon et al. (2000) use a dataset constructed by Kraay et al. (2005) to test a
portfolio model on a set of developing and industrial countries. Gourinchas and
Rey (2007) use monthly data and an intertemporal budget constraint view to
measure external imbalances in the United States. External imbalances should
be especially disruptive in developing markets. IMF (2005) uses a methodology
close to Gourinchas and Rey (2007) to show the different roles played by
valuation effects in emerging and industrial countries. The idea behind this is that
valuation effects are destabilizing in developing countries because of liability
dollarization (see for example Obstfeld, 2004). The more an economy is
dollarized, the worse will be a reaction of its net foreign assets position to
depreciation. And since the reaction of the exchange rate to excess external
liabilities will be to depreciate, a dollarized indebted country should become
even more indebted, unless it runs substantial trade surpluses. On the other
hand, if these surpluses are not accompanied by a surge in productivity, it should
take place thanks to a shift in demand from tradables to nontradables, and this is
possible only through further real exchange rate depreciation. This mechanism
initiates a vicious circle that badly affects firms' balance sheets and their capacity
to invest, thus leading to an output collapse. Indeed, IMF (2005) finds that
valuation effects play a stabilizing role in industrial countries but not in
developing countries. Although stock imbalance measures proved useful at
predicting future flows (see Lane and Milesi-Ferretti, 2001, IMF, 2005 and
Gourinchas and Rey, 2007), no attempt has been made at using them to predict
the particular phenomenon of sudden stops in capital flows, which can be
defined as a sharp, disruptive reversal in the current account.
The mean reversion property of current account has several implications for
international macroeconomics. First, a stationary current account is consistent
with sustainability of the external debts. In this case, there is no incentive for the
— 17 — government to make drastic policy changes and default on its international debts
in the near future. Second, stationarity of the current account validates the
modern intertemporal model as, theoretically, the model combines the
assumptions of perfect capital mobility and consumption smoothing behavior to
postulate that the current account acts as a buffer to smoothing consumption in
the event of shocks. From an empirical point of view, the stationarity and
sustainability of OECD current account balances has been the focus of many
researchers over a number of years6. The literature on the sustainability of the
current account examines the question within two alternative frameworks.
On the one hand, a time series perspective is employed where researchers
investigate either the long-run relationship between exports and imports or the
stationarity of the external debt process (see Chortareas et al., 2004)7. With the
exception of Liu and Tanner (1996), who consider the impact of structural
breaks, the above mentioned studies generally find that current accounts are
non-stationary for OECD countries. Tests that rely on linear approximations are
likely to be imprecise on short samples when the observed current account is
persistent, as it typically is. A persistent current account does not necessarily
mean a non-stationary one. A stationary current account will be considered
persistent if its process of mean reversion is slow. The small sample problems
can occur even if the current account is stationary but persistent. Therefore,
panel data can improve the information in relatively short sample databases.
On the other hand, panel unit root techniques have been employed since
unit root tests applied to single series suffer from low power. In recent years a
number of alternative procedures have been proposed to test for the presence
of unit roots in panels that combine the information from the time series
dimension with that from the cross-section dimension. Studies that employ
panel data methods include Wu (2000), Wu et al (2001), Holmes (2006) using
Im, Pesaran and Shin (2003) test (IPS) and cointegration tests. However, due to
the heterogeneous nature of the alternative hypothesis in their test, one needs
to be careful when interpreting the results, because the null hypothesis that
there is a unit root in each cross section may be rejected when only a fraction of
the series in the panel is stationary. Moreover, the hypothesis that the current
account balances or the debt series adjust to long-run equilibrium in a
continuous fashion is troublesome when we suspect that in many cases
discontinuities may be present in the mean-reversion process. We can identify
several sources of breaks rooted in policy or institutional investors' behavior.
6
See, inter alia, Trehan and Walsh (1991), Otto (1992), Wickens and Uctum (1993), Liu and
Tanner (1996), Wu (2000), Wu et al. (2001), Holmes (2006) and Holmes et al. (2007).
7
The strand of this empirical literature using single equation unit root tests usually rejects the
mean reverting behavior of the current account. See, among others, Husted (1992), Ghosh
(1995), or Bergin and Sheffrin (2000).
— 18 — Instituto de Estudios Fiscales
The presence of high government debt may have repeatedly induced abrupt
corrective actions requiring sudden adjustments. More specifically, the fiscal
health of the government tends to affect international investors' perception of
expected profitability and the investment climate in the country. This
perception, in turn, may trigger abrupt reversals in capital flows and imbalances
in the current account (i. e. the EMS crisis in the early 1990s). Another channel
that may lead to discontinuities in the series is the level of a country's
indebtedness, which reflects the willingness of foreign lenders to hold domestic
assets. Investors may be unwilling to lend beyond a level of foreign debt they
consider “normal” and withdraw large amounts of funds, creating majors
imbalances in the balance of payments.
4. THEORETICAL FRAMEWORK
4.1. The classical flow equilibrium approach: sustainability of the
current account and the intertemporal budget constraint
According to Taylor (2002) sustainability of the current account can be
defined as the ability of an economy to satisfy its long-run intertemporal budget
constraint without a drastic change in private sector behavior or policy shifts. It
views the current account as the equilibrium outcome of forward-looking saving
and investment decisions taken by rational individuals and driven by
expectations of productivity growth, government expending, interest rates and
other factors. This view emphasizes the role of the current account as a buffer
against transitory shocks in productivity or demand in order to smooth the
intertemporally-optimal consumption path. As we previously claimed, this is a
rather general concept and does not depend on any particular model, with the
advantage of its easy testability. According to Trehan and Walsh (1991), current
account stationarity is a sufficient condition for the intertemporal budget
constraint to hold.
Consider a stochastic model with zero growth. The one period budget
constraint is,
C t + It + Gt + NFA t = Yt + (1+ ii )NFA t −1,
(1)
where Ct ,It ,Gt,NFA t and Yt are consumption, investment, government
consumption, net stock of debt and income respectively. it is the world interest
rate. Rearranging (1) and from national accounts identities we have that,
NFA t = (1+ i i )NFA t −1 + NX t ,
— 19 — (2)
where NX t is the net exports. Iterating (2) forward and assuming that the
expected value E(i t | ϕ t−1 ) = i , with ϕ t−1 being the information set available in
t − 1, we get
∞
j
T
⎛ 1 ⎞
⎛ 1 ⎞
NFA =
⎟ E(NFA t +T | ϕt−1).
⎜
⎟ E(NX t + j | ϕt −1) + lim ⎜
T→∞⎝ 1+ r ⎠
1+
r
⎝
⎠
j=0
∑
(3)
Equation (3) simply states that international agents are able to lend to an
economy if they expect that the present value of the future stream of next
exports surpluses equals the current stock of foreign debt. Hence, the
sustainability hypothesis, or long run budget constraint implies that:
T
⎛ 1 ⎞
lim ⎜
⎟ E(NFA t+T | ϕ t−1 ) = 0
t→∞⎝ 1+ i ⎠
(4)
This transversality condition means that the present value of the expected
stock of debt when t tends to infinity must equal zero, that is, a no-Ponzi game
condition. Following Trehan and Walsh (1991), given that the current account
CA t = NFA t − NFA t−1 , a sufficient condition for (4) to hold is that the current
account is an I(0) stationary process. In the more realistic case of an economy
with a positive rate of growth of output, we have that the sustainability condition
t
holds if the ratio ca t = CA
is I(0) stationary. This means that sustainability is
Y
t
possible with current account deficits as far as they do not grow faster than
output in expected value.
An obvious test of sustainability is hence a unit root test on ca t . This is what
most of the literature has previously used as a test of sustainability. However,
note that we are dealing here with expected values of future events. Changes
in the agents' perceptions about risk, portfolio allocation decisions, future
policy changes, transaction costs in international financial flows, among others,
can lead to changes in the dynamics of current account mean reversion and,
hence, equilibrium values of the current account. As previously mentioned,
Taylor (2002) sees the speed of convergence towards equilibrium as a
summary statistic of the degree of capital mobility. This is because it reflects
how agents are prepared to allow for periods of current account deficits
(surpluses) above the perceived equilibrium value. If, given the international
financial environment, agent's perceptions about, for instance, the relative risk
of US denominated assets changes due to large observed current account
deficits, the speed of mean reversion and the mean of the current account
itself would also change. That is, changes in the current account affecting the
agent's perception can trigger adjustment dynamics leading to discontinuities in
the time series. In this sense, it may be the case that tests that do not consider
the existence of breaks are misspecified and reach wrong conclusions about
the sustainability of the current account or arrive at too simplistic descriptions
— 20 — Instituto de Estudios Fiscales
of the current account dynamics. Moreover, the special nature of the financial
markets, characterized by contagion effects may give rise to sudden stops or
even reversals in the asset holdings leading again to breaks in the time series
and to the existence of cross-section dependence. This fact may again lead to
misleading conclusions. In this research we overcome these two problems
through a new panel unit root test that considers the existence of multiple
breaks and cross-section dependence.
Although the saving-investment equilibrium approach does provide an
analytical basis for the evaluation of external positions, its almost exclusive
concern with flows limits its ability to assess the viability and adequacy of
external indebtedness, a stock problem by nature.
4.2. The stock approach: the arithmetic of intertemporal solvency
(Net international debt to GDP ratio)
The arithmetic of solvency starts from the notion that an economy is
intertemporally solvent if its (net) foreign indebtedness is no larger than the
present discounted value of the stream of its future non-interest surpluses. The
practical difficulty with this approach is that in principle any level of external
debt is consistent with solvency provided that sufficient trade surpluses are
generated in the indefinite future (Milesi-Ferretti and Razin, 1996). Thus, to
make this approach operational, researchers typically assume that the economy
targets a given debt-to-GDP ratio, and consider the particular case in which
current policy would remained unchanged into the indefinite future (Corsetti
and Roubini, 1991). The arithmetic of solvency is primarily concerned with the
question of whether net external liabilities grow less rapidly than their (marginal)
rate of return so that the present discounted value of net liabilities converges to
some finite quantity. In practical terms, the arithmetic of solvency examines
whether the net debt/GDP ratio grows more or less rapidly than the difference
between the real interest rate and the economy's growth rate. Following
Chortareas et al (2004) let us start with a stylized version of the nominal balance
of payments identity defined in domestic currency:
NFA t = ER t ∆L t − ∆A t ≈ NX t + i∗t ER t L t−1 − i t A t−1
(5)
where NFA t is the net foreign assets position, NX t stands for trade balance,
A(L) are domestic (foreign) assets held by foreigners (domestic residents), i (i∗ )
are the nominal rates of return on domestic (foreign) assets, and ER is the
domestic price of the foreign exchange rate. Deflating by nominal GDP, and
regrouping terms, the former identity can be rewritten as:
∆nfa t ≈ c t + ~
rt nfa t−1
— 21 — (6)
where
c t = τ t + (i t − i∗t − e t )b ∗t−1
is the primary current account deficit,
b t − b ∗t
is net foreign indebtedness, τ t is net exports, and ~rt = it − p t − y t is
the growth-adjusted real return on net foreign debt. Further, e = ∆ logER t ,
p = ∆ logPt , and y t = ∆ log Yt , all other lower case letters denote variables as a
ratio to nominal GDP. If (6) is deflated by a price index, f and c are real foreign
debt and current account, and ~r is the real interest rate. Assuming ~r > 0, solving
(6) forward, and imposing the no-Ponzi game condition, the Intertemporal
Budget Constraint (IBC) is:
nfa t =
n
nfa t = −
∑ ρ t c t+i
(7)
i=1
(1 + ~rt+s )−1 If this conditions holds, current and future discounted
with ρ t = Π ns=1
primary current account surpluses are sufficient to pay off initial indebtedness.
The traditional sustainability approach tests for the stationarity on nfa t . In the
present exercise we take account of the valuation effects of stocks of foreign
assets and liabilities using the new External Wealth of Nations Mark II (EWN II)
database provided by Milesi-Ferretti and Lane (2007). According to them the
size of countries' external portfolios is now such that fluctuations in exchange
rates and asset prices cause very significant reallocations of wealth across
countries, playing the exchange rate a dual role influencing both net capital flows
and net capital gains on external holdings.
4.3. The unified approach of Gourinchas and Rey (2007): foreign debt
and the current account
In this subsection we summarize the model developed by Gourinchas and
Rey (2007). This model follows an intertemporal approach and is based on two
elements: an intertemporal budget constraint and a long run stability condition.
They start from a country's intertemporal budget constraint and derive two
implications. The first one is a link between the net foreign asset position and
the future dynamics of the current account. If total returns on NFA are
expected to be constant, today's net foreign liabilities must be offset by future
trade surpluses (the so called “trade channel”). However, in the presence of
stochastic asset returns, the expected capital gains and losses on gross external
positions constitute a complementary adjustment tool called the “valuation
channel”. The external constraint implies that today's imbalances must predict
either future changes in the trade balance (flow adjustment), future movements
in the returns of the NFA portfolio (changes in the stock of foreign assets), or
both. In the short and medium term, most of the adjustment goes through asset
returns, whereas at longer horizons it occurs via the trade balance.
— 22 — Instituto de Estudios Fiscales
The value of assets owned by domestic residents held abroad (A) minus the
value of domestic liabilities to the rest of the world (L) is called the national net
foreign asset position (NFA) . If its net foreign asset position is positive
(NFA > 0) , the country is a net creditor to the rest of the world. Conversely, if
NFA is negative (NFA < 0) then the country is a net debtor, because its
outstanding liabilities to the rest of the world exceed its claims on the rest of the
world. All nations are subject to a budget constraint that requires that the value
of gross domestic expenditure, (GDE) or absorption, plus the change in the
stock of foreign assets owned by domestic residents (A t − A t−1 ) equals the value
of gross domestic product (GDP) plus the change in the stock of domestic debt
owed to foreigners (L t − L t −1 ) . Combining this relationship with the definition of
the current account, it follows that the change in the net foreign asset position is
the same as the balance on the current account.
(GDPt − GDE t ) + NFIt + UTt = (A t − L t ) − (A t−1 − L t−1 )
(8)
Substituting in the definition of the net export balance (NX t = GDPt − GDE t )
and net foreign asset position (NFA t = A t − L t ) , this simplifies to:
NX t + NFIt + UTt = CA t = NFA t − NFA t−1
(9)
which says that the change in the net foreign asset position is the sum of net
exports, net foreign income, and unilateral transfers or the balance on the
current account.
Therefore, if the current account is in deficit (CA < 0) , the change in the net
foreign asset position is negative, indicating that the increase in foreign debt was
greater than the increase in foreign assets over the year. A negative change in
the net foreign asset position is referred to as a net capital inflow, since more
capital flowed into the country through additions to the level of foreign debt
than flowed out through purchases of foreign assets. Future current account and
net foreign asset positions are related to the present current account and net
foreign asset positions through future net foreign income flows8. The extent of
these flows is influenced by the rates of return on foreign assets and foreign
debt. Net foreign income is essentially the difference between interest earned
on foreign assets and interest paid on foreign liabilities:
NFIt = r A A t−1 − r L L t−1
(10)
where r A is the rate of interest residents earn on their foreign assets and r L is
the rate of interest that the country pays on its foreign liabilities. Theoretical
8
See Lane and Milesi-Ferretti (2002) and Gourinchas and Rey (2007) for a more complete
discussion of the longer term relationship between the US net exports deficit and
revaluations of the US net foreign asset position.
— 23 — analyses typically assume that there is no differential between the interest rate
on foreign assets and debt, and that the interest rate on foreign debt exceeds
the growth rate of nominal GDP, which suggests that the economy must shift to
a net export surplus to maintain its current negative net foreign asset position.
In textbook examples there is no distinction between r A and r L , because they
assume there is only one traded asset. However, this assumption is far from
reality9, so it is important to allow for differences between r A and r L :
NFIt = (r A − r L )A t−1 + r L (A t−1 − L t−1 ) = (r A − r L )A t−1 + r L NFA t−1 (11)
Substituting expression (11) into (9) above, it follows that:
NX t = NFA t − (1 + r L )NFA t−1 − (r A − r L )A t−1 − UTt
(12)
Dividing through by the level of GDP and imposing the foreign debt
sustainability condition that the ratio of NFA to GDP be constant at nfa∗ , we
find that the critical net exports to GDP ratio, nx ∗ at the current gross foreign
asset to GDP ratio a∗ and typical unilateral transfer to GDP ratio ut ∗ is:
nx ∗t = (g − r L )nfa ∗ − (r A − r L )a ∗ − ut ∗
(13)
where g is the growth rate of nominal GDP.
According to Kouparitsas (2005) when economists want to assess sustainability
of the current account, they begin by calculating the net exports to GDP ratio
that would be required to maintain the current net foreign assets to GDP
ratio, nfa∗ . He refers to this as the critical net exports to GDP ratio, nx ∗ . Net
exports to GDP ratios above nx ∗ will increase the nation's net foreign assets to
GDP ratio above nfa∗ , while net exports to GDP ratios below nx ∗ will decrease
it. For reasons explained below, negative net foreign asset positions are typically
associated with a positive nx ∗ . Another way of stating this is that a country must
give up a fraction of all future GDP equal to nx ∗ to maintain its current negative
net foreign asset position. A country's current net foreign asset position is
considered unsustainable if the associated nx ∗ is a relatively large fraction of
GDP. Similarly, a current account deficit is considered unsustainable if it
maintains or leads to an unsustainable net foreign asset position. According to
this analysis, nx ∗ depends not only on nfa∗ , which is weighted by the difference
between the growth rate of nominal GDP and the interest rate on foreign debt,
9
The experience is inconsistent with standard theoretical assumptions in many countries, like
the US. First, the return the US earns on its private foreign assets exceeds the rate it pays on
its private foreign debt. Second, on average, rates of return on most classes of US foreign
debt have been roughly equal to the growth rate of nominal GDP.
— 24 — Instituto de Estudios Fiscales
but also the current ratio of domestic gross foreign assets to GDP, a∗ , which is
weighted by the difference between the interest rates on foreign debt and
foreign assets, and the typical ratio of unilateral transfers to GDP, ut ∗ .
A by-product of this analysis is the current account to GDP ratio, ca ∗ , that
would be required to maintain nfa∗ . ca ∗ only depends on nfa∗ , which is
weighted by the growth rate of nominal GDP. Through a similar analysis, one
can show that the critical current account to GDP ratio ca ∗ is:
ca ∗ = g ⋅ nfa ∗
(14)
Moreover, many statistic databases do not take into account the unrealized
capital gains from both changes in local currency prices and exchange rate
adjustments and this mechanism can be of increasing importance in a financially
integrated world10.
Although the theoretical relation (14) should hold between the current
account and the net foreign assets, Gourinchas and Rey (2007) argue that they
use trade balance instead of the current account to avoid possible discrepancies
in the valuation of capital gains11. Consequently, they consider the accumulation
identity for net foreign assets between t and t + 1 :
NFA t+1 = R t+1(NFA t + NX t ),
(15)
where NX t are the net exports (difference between exports, X t , and imports,
Mt ) and NFA t are net foreign assets (difference between gross foreign assets,
A t , and gross foreign liabilities, L t , measured in the domestic currency).
According to equation (15), the net foreign position would increase with net
exports and with the total return on the net foreign asset portfolio R t+1 (see
equation (2) above) .
Next, the model is log-linearized. The following assumptions should hold:
Assumption 1: (a) The ratios A t / W t , L t / W t , X t / W t and M t / W t are
stationary, where Wt represents total household wealth. (b)
The steady state values of the ratios, denoted µ aw , µ lw , µ xw
and µ mw respectively, satisfy µ aw ≠ µ lw and µ xw ≠ µ mw .
Assumption 2: The growth rate of household wealth Wt+1 / Wt is stationary
with steady state value γ .
10
See Lane and Milesi-Ferretti, (2001) and Tille (2003) for detailed discussion of the size and
history of valuation effects for the US and other nations.
11
However, in order to ease the comparison between the three theoretical approaches that
we present in this research, in the empirical application we substitute CA t by NX t in equation
(15).
— 25 — Assumption 3: The return to the net foreign asset portfolio R t is stationary
with a steady state value R that satisfies γ < R .
Concerning Assumption 1, this is not particularly restrictive. The first part of
the assumption would be verified in any model where exports, imports,
external assets, liabilities and household wealth grow at the same rate along a
balanced growth path. This will be the case in a wide variety of models, as long
as assets and liabilities are not perfect substitutes. The second part of the
assumption guarantees that some ratios are well defined. Assumption 2 is also
an implication of the existence of a well-defined balanced growth path.
Assumption 3 has an intuitive interpretation in this context: manipulating
equation (15) if Assumption 3 holds, the steady state ratio of net exports to net
foreign assets is stationary with an unconditional mean NX / NFA that satisfies:
NX
= ρ − 1< 0,
NFA
(16)
where ρ = γ / R < 1 implies that the real growth rate of wealth is lower than the
rate of return of the net foreign asset portfolio. Therefore, countries with
steady state creditor positions (NFA > 0) should run trade deficits (NX < 0) ,
whereas countries with steady state debtor positions (NFA < 0) should run trade
surpluses (NX > 0) . Lane and Milesi-Ferretti (2002) point out that the correlation
between the change in the net foreign asset position at market value and the
current account is low or even negative. They also note that rates of return on
the net foreign asset position and the trade balance tend to commove
negatively, suggesting that wealth transfers affect net exports. Moreover, Lane
and Milesi-Ferretti (2004) show exchange rate effects on rates of return of
foreign assets and liabilities.
5. ECONOMETRIC METHODOLOGY AND RESULTS
In this section we describe the testing strategy we use to address the
theoretical issues described above. The empirical application is based on a panel
database that consists of 20 OECD countries, both European and from the rest
of the world. The sample covers the period 1970-2006, and the data has been
obtained from the World Bank and the new External Wealth of Nations Mark II
(EWN II) database provided by Milesi-Ferretti and Lane (2007a). The two
variables of interest are the current account balance as a percentage of GDP
( ca i, t ), and the net foreign assets stock also as a percentage of GDP ( nfa i,t ). We
first test, using panel methods, both for the sustainability of the current account
and the solvency of the net foreign assets position of our group of countries.
Then, following what we call the “unified approach” by Gourinchas and Rey
— 26 — Instituto de Estudios Fiscales
(2007), we study the stability and the sign of the relationship linking the two
variables.
Concerning the two first hypotheses, we have applied panel data based test
statistics following a two-step testing strategy that addresses the problems
related to the issues of multiple structural breaks and cross-section
dependence12.
First, we have tested for the sustainability of the current account and for the
external solvency by allowing for multiple structural changes in a panel setting
that, to the best of our knowledge, has not been applied yet in this literature.
Previous evidence has revealed that there might be some events that affect the
current account and the net foreign asset position in a permanent way. It is well
known that non accounting for structural breaks biases both unit root and
stationarity tests towards concluding in favor of non-stationarity in variance13.
Thus, this feature should be of special interest in our case, since this type of
variables may be affected by major events such as currency crises or economic
integration processes during the analyzed period. Second, we consider the
existence of cross-section dependence amongst the individuals in the panel.
Cross-section independence is hardly found in practice, especially when using
macroeconomic time series that derive from globalized financial markets, as it is
the present case. Moreover, it is worth mentioning that the existing literature
has evidenced an increase in the degree of market integration, which should
lead to higher correlation between financial and macroeconomic aggregates at
the international level. As panel data unit root and stationarity tests are known
to be biased towards concluding in favor of variance stationarity when
individuals are cross-section dependent –see O'Connell (1998) and Banerjee,
Marcellino and Osbat (2004, 2005)– the issue of cross-section dependence is of
great importance. Therefore, we suggest computing the test statistic in Pesaran
(2004) and Ng (2006) to assess whether the individuals in the panel are cross­
section independent. Furthermore, Ng's (2006) statistic is quite convenient
since, in addition to testing for the null hypothesis of cross-section
independence, it provides guidance about the best way to model cross-section
dependence.
The application of this statistic reveals that cross-section dependence is
present in the panel data sets that we study. Then, our analysis considers two
different ways to accommodate cross-section dependence. First, following the
approach by Carrion-i-Silvestre et al. (2005) we compute the bootstrap critical
values of the panel data stationarity test statistic, which allows us to consider a
12
We have applied as well classical panel unit root and stationarity tests without structural
breaks finding mixed results. These results are available upon request from the authors.
13
See Perron (1989) for univariate statistics, or Carrion-i-Silvestre, del Barrio and LópezBazo (2001) for panel data statistics.
— 27 — wide form of cross-section dependence. Second, we compute the panel data
unit root and stationarity test statistics proposed in Harris et al. (2005) and Bai
and Carrion-i-Silvestre (2009), which model the presence of cross-section
dependence through the estimation of approximate common factor models as
in Bai and Ng (2004). In both cases, the analysis considers the existence of
multiple structural breaks. In addition, the approach that is adopted here is
general enough to consider the non-break situation as a particular case
embedded in the testing procedure. Therefore, our analysis does not impose
the existence of structural breaks, but accounts for the possibility that they are
present in the data.
Finally, note that proceeding in this fashion accounts for the existence of a
tension or trade-off between cross-section dependence and misspecification
concerning the presence of structural breaks: the former introduces a bias towards
stationarity in variance while the bias due to the latter goes in the opposite
direction. This feature implies that the empirical analysis of the current account
balances should be addressed carefully to avoid the effects of this tension.
Then, we analyze the relationship between the current account balance
(cai, t ) and the stock of net foreign assets (nfai, t ). Once the panel methods have
concluded that the two variables are stationary, we can use the Bai-Perron
procedure to assess the stability and the sign of the relation linking them. This
will constitute a test of the unified approach of Gourinchas and Rey (2007) as
described above. Its empirical implementation on a large cross-country time
series sample poses two main issues. First, the model defines a long-run
relationship between the ratio of CA and NFA. However, given the imperfections
in international financial and factor markets, stock equilibrium does not hold in
every point in time but is achieved gradually in the long-run. Therefore, in the
empirical analysis, this relation might also present instabilities. Second, it seems
reasonable to assume that countries can differ in the market imperfections and
barriers to portfolio reallocation that govern the short term dynamics, and
perhaps even in the parameters characterizing the long-run equilibrium. Thus,
we must take into account the very likely possibility of parameter heterogeneity
across countries.
In the third part of this paper we first propose to apply in a single-country
context the methodology developed by Bai and Perron (1998) to ascertain
possible breaks in the relationship. Additionally, we use standard VAR methods
of estimation to capture the dynamics of the adjustment. A salient feature of our
analysis is that we do not use detrended variables as in Gourinchas and Rey
(2007). We think that this would neglect relevant information that is embedded
in the trend of the variables. As Gourinchas and Rey (2007) claim, trends might
be representing structural changes in the world economy, such as financial and
trade globalization. Therefore, we have considered an econometric strategy
— 28 — Instituto de Estudios Fiscales
that allows identifying and including the possible structural breaks in the
estimation of the relationships between the variables.
5.1. Testing for current account sustainability and external debt
solvency: panel analysis
We use panel data methods in the empirical analysis of the theoretical
hypotheses described in subsections 4.1. and 4.2. above. For simplicity, we
present the testing strategy and the results simultaneously, as the same tests and
procedures are applied in the two cases. However, in the discussion, we will
explain separately the main conclusions.
5.1.1. Testing for the presence of multiple structural breaks
The first stage of our analysis consists of assessing the presence of structural
breaks affecting the ca i, t and nfa i,t time series using the following specification:
y i,t = α i +
mi
∑ θi,k DUi,k,t + e i,t ,
(17)
k =1
where y it is the variable of interest, whereas t = 1,K,T , i = 1,K,N , with
DUi,k, t = 1 for t > Tbi,k and 0 elsewhere – Tbi,k denotes the k th break point for the
{ }
i th individual, k = 1,K,m i – and where e i,t
are assumed to be a stationary
process satisfying the strong-mixing conditions given in Phillips (1987) and
Phillips and Perron (1988).
This specification permits a high degree of heterogeneity assuming that the
structural breaks may have different effects on each individual time series. For
this purpose, the break points are located at different dates for each individual,
and the individuals may have different number of structural breaks. Under these
conditions, the estimation of the number and position of the structural breaks, if
any, can be carried out using the sequential testing procedure proposed by Bai
and Perron (1998). When computing the statistic we have to specify a maximum
number of structural breaks, which in this case has been set equal to m i = 5 ∀i .
The number of structural breaks is estimated using critical values at the 5%
level of significance. It is worth mentioning that the application of the Bai-Perron
methodology to estimate the number and position of the structural breaks
requires the variables under analysis to be stationary in variance, which is
consistent with the null hypothesis that we have specified, i.e., that the solvency
hypothesis holds. Furthermore, the test statistic that is used is consistent against
the alternative hypothesis of non-stationarity in variance, even when structural
breaks are present in the analysis – see Lee, Huang and Shin (1997), Kurozumi
(2002) and, Carrion-i-Silvestre (2003), among others.
— 29 — Panel A in Table 1 reports the estimated number and position of the
structural breaks for each individual in the current account panel data set. We
can see that, except for Italy and New Zealand, the procedure detects at least
one structural break for each time series, which indicates that previous
analyses in the literature that do not account for the presence of structural
breaks may have led to misleading conclusions. It should be stressed that the
estimated number of structural breaks does not attain the maximum that has
been defined.
Concerning nfa i,t , we present the breaks and their position in Panel A of 4.
With the exception of Germany, we find at least one structural break in all
the countries in our group. We should note that the two variables we
consider, current account balance and nfa i,t , are very different in nature:
whereas the first one is a net flow, the second one, nfa i,t is a cumulated series
(a stock). Therefore, we do not expect to find that the breaks are placed at
the same dates for the two variables. As the literature on current account
reversals describes (see Freund (2005) and Debelle and Galati (2007), for
example), the adjustment usually takes place once the current account
imbalance reaches a certain threshold. The effect on the net foreign asset
position would critically depend on the relative position of assets and
liabilities when the event takes place.
Figure 8 depicts the CA time series for all the countries involved in our
analysis along with the estimated deterministic component. The countries have
been divided according to their condition of EU members during the studied
period. This presentation allows us to establish a comparison of the break dates
and the direction of the changes that have been estimated. In Table 2 we
present an approximation to the main events explaining the structural breaks
found in the data. We have ordered the countries following two criteria: (i) their
EMU (or EMS) membership and (ii) their external position in terms of the
current account. Two countries, Ireland and France could not be clearly placed
in a current account category and, hence, they are considered separately. In the
Table we have limited ourselves to the main milestones in European integration
and international economic events. Other issues, however, may explain a
particular structural break. We next analyze the countries individually.
At the beginning of the 70's, the first oil shock triggered the collapse of the
Bretton Woods system inducing effects on different countries. Belgium and
Austria, decided to link its currency to the Deutsche Mark at the end of Bretton
Woods – therefore, a policy change may have happened in 1974 and 1975 for
Belgium and Austria, respectively.
Two non-EMU countries suffered structural changes at the beginning of the
eighties. Australia in 1980, when the dollar experienced a depreciation linked to
a terms of trade worsening – in 1979 the Australian financial market
— 30 — Instituto de Estudios Fiscales
experienced a process of deregulation, and the dollar freely floated in 1983. The
break in Norway in 1979 is possibly linked to the increase in oil prices.
A large group of countries have a structural break in the mid-eighties. Both
Belgium and Germany followed recovery programs. For example, president
Martens in Belgium devalued the frank in 1982 and started an export-led policy.
Ireland also devalued in 1983 in an answer to a twin deficits problem, followed
by a tight fiscal policy14. Austria in 1980 started a system of cooperative
arrangement for its exchange rate. Finally, Portugal suffered a deep recession,
with terms of trade losses, fiscal deficits and increase in foreign debt service.
Concerning non-EMU countries, the Reagan administration started a
program at the beginning of the eighties that reduced policy intervention and
allowed the free floating of the dollar. In early 1981, the new Reagan
Administration decided to move away from what it judged to have been the
heavy intervention inherited from the previous administration. From 1981
through early 1985, the dollar continued to strengthen, for several reasons. US
monetary conditions were restrictive in the context of a robust recovery, and
prospects for continued large US fiscal deficits exerted upward pressure on real
interest rates. Meanwhile, monetary authorities abroad initially were reluctant
to raise interest rates because their recoveries appeared more fragile.
Investment, including foreign investment, boomed in the United States,
attracted by the increasingly favorable business climate. In addition, dollar­
denominated assets were sought as a safe haven following the onset of the
international debt crisis and amid apprehensions about the political situations in
some European countries.
Another large group of structural changes is found during the first half of the
nineties. Most of the breaks are linked to the free capital movements in Europe
and the German Unification in 1990, together with the EMS crises in 1992 and
1993. Portugal and France suffered a slowdown in economic activity in an effort
to fulfill the Maastricht criteria. In the case of Austria, EU membership occurred
in 1995, together with Sweden and Finland. The only structural break that
Finland suffered occurred in 1994, the year of the referendum for EU accession.
Sweden presents two structural breaks: the first one (in 1994) can be related to
inflation targeting policy that started in 1993, whereas the second one (in 2001)
is placed at the peak of an economic expansion.
Finally, the end of the nineties and the beginning of 2000 accumulates
another group of structural changes. Those in EMU countries and the US are
14
Membership of the EMS always posed problems for Ireland by virtue of the fact that the
UK, the country's major trading partner, is not a member of the system. Such problems
became most acute when a depreciation in Sterling put pressure on Irish companies in
traditional industrial sectors. Such considerations prompted a devaluation of the Irish pound
at the March 1983 re-alignment.
— 31 — linked to the creation of the monetary union in 1999, the launching of the euro
in 2001 and its effects on the dollar15. At the same time, Norway established an
inflation targeting strategy, whereas Sweden, also outside the EMU,
experienced an economic expansion. In contrast, the Asian crisis affected the
demand of commodities and deteriorated Canadian dollar (and its terms of
trade, suffering an adverse current account shock). Beginning in the summer of
1997, Malaysia, Indonesia, Thailand and South Korea (and some other Asian
countries) fell into a serious recession, sparked by the collapse of their pegged­
exchange-rate regimes. As these countries are large users of raw materials, their
recessions led to a significant fall in the world's demand for raw materials, and
thus a large decline in raw materials prices. In the next year or so, the average
prices of raw materials fell by about 30 per cent. All countries that export raw
materials experienced a sudden decline in demand for their currencies, which
lost value as a result – Canada, New Zealand, and Australia. This type of shock
is a negative current-account shock, because it reflects a reduction in the
demand for Canadian goods or services, the transactions of which are recorded
in the country's current account of the balance of payments. In Japan, the real
estate bubble burst and the current account was declared to be a monetary
policy target.
From the joint analysis of Table 4 and Figure 9, a similar pattern can be found
in the structural breaks of nfa i,t . In Table 4 the breaks are classified according to
the type of country (EMU member or third country) and its NFA position
(positive or negative). In addition, an arrow indicates the direction of the break
(where ↑ stands for an improvement in the position and ↓ for a worsening).
With two exceptions (Australia in 1978 and Belgium in 1977), the rest of the
breaks are accumulated in the mid-eighties and the end of the nineties-2000.
The economic and political events described above are also valid for the NFA
variable and describe the reasons for the occurrence of the breaks.
We have compared the structural breaks found in the current account
balance over GDP with the results of a strand of literature that studies the
current account reversals16. The main conclusion that can be derived is that the
structural changes detected in the variable coincide with reversals or
15
The dollar broadly strengthened against other currencies after the mid-1990s because
market participants expected to receive higher rates of return on their investments in the US
than abroad. For example, consider for a moment the fate of the euro versus the dollar since
the euro's launch on january 1, 1999. The dollar strengthened by 30% against the euro
primarily because market participants anticipated brighter prospects and higher rates of
return in the US than in Euroland, and capital flowed out of euro-denominated assets into
equities, bonds and other US investments.
16
See, for example, Freund (2005), Debelle and Galati (2007) and De Hann et al (2008) for a
detailed list of these episodes.
— 32 — Instituto de Estudios Fiscales
adjustments17 in the majority of the countries. This is, for example, the case of
Australia in 1980, of Canada, Finland and Sweden in the mid-nineties, and
Portugal, the UK and the US in the eighties. However, we have identified slightly
different dates for the structural breaks in nfa i,t . This result is not surprising, as
this variable is a stock and a process of adjustment takes several years. Therefore,
the bulk of the structural changes are found in the eighties and the nineties, but
also around 2000. For example, whereas the Finish current account has a break in
1994, its net foreign assets have two breaks: one in 1996 and another one in
2001. In the US, the dates are 1982 for the current account and 1984 in nfa i,t
Later, a second break appears in 1999 in the current account, coincident with a
break in net foreign assets. The literature dates a reversal in 2000.
5.1.2. Testing I(0) stationarity on individual time series
Once the break points have been dated, we proceed to analyze the order of
integration of the yt time series. The estimation of the model in (17) with the
break points that have been obtained above can be used to compute the individual
stationarity test in Kwiatkowski et al. (1992) –henceforth, KPSS statistics– given by
η̂i (λ i ) =
ω̂i−2 T−2
T
∑ Ŝi,t2 ,
(18)
t=1
where Ŝ i,t = ∑tj=1 ê i, j is the partial sum process that is obtained using the
estimated OLS residuals of (17), ω̂i2 denotes a consistent estimate of the long­
run variance of the error term e i,t , which, based on the evidence reported in
Carrion-i-Silvestre and Sansó (2006), has been estimated following the
procedure described by Sul et al. (2005), using the Quadratic spectral kernel. In
′
i
i
/ T,...,Tb,m
/ T ⎟⎞ , which
(18), λ i is defined as the vector λ i = (λ i,1,...,λ i,mi )′ = ⎜⎛ Tb,1
⎝
i, j
⎠
indicates the relative position of the dates of the breaks on the entire time
period T for each individual. Thus, the computation of the individual KPSS
statistic permits to get a first analysis of the stochastic properties current
account and the net foreign asset position (see Panel A in Tables 1 and 4,
respectively). The statistics in Panel A offer the computation of the individual
KPSS along with the corresponding simulated critical values at the 5 and 10%
level of significance. Focusing on the individual statistics of ca i, t , we can see that
17
A current account adjustment is defined by three conditions: (i) the current account should
exceed 2% of GDP prior to the adjustment; (ii) the average deficit should decline by at least
2% of GDP over three years and be reduced by at least a third; (iii) the largest deficit during
the five years after the peak should not be wider than the smallest deficit during the three
years before the peak (Debelle and Galati, 2007).
— 33 — the null hypothesis of I(0) cannot be rejected at the 5% level of significance for
fifteen out of twenty countries – the exceptions are Ireland, Japan, Netherlands,
Portugal, and Sweden. Therefore, the constraint is met for the majority of the
countries in the panel data set, although the fact that for these countries ca i, t is
found to be I(0) evolving around a broken deterministic component implies that
the current account is not sustainable. The results are similar in the case of nfa it :
the null hypothesis of stationarity is rejected at 5% in eight cases (Austria,
Belgium, Italy, Japan, Spain, Sweden, UK and US).
This individual based inference can be improved if we combine the individual
statistics through the definition of panel data statistics. Thus, the literature on
non-stationary panel data statistics argues that a better characterization of the
stochastic properties of the time series can be obtained if we increase the
amount of information when performing the inference. However, some cautions
have to be taken when computing these panel-data-based statistics, since some of
them rely on the critical assumption of cross-section independence. This
assumption is investigated in the next section for our panel data sets.
5.1.3. The issue of cross-section independence
The independence assumption imposed in the so-called first generation panel
data statistics has been widely criticized in the recent literature, since it has been
shown that non accounting for cross-section dependence amongst the individuals
might bias the statistical inference in favor of variance stationarity – see Banerjee
et al. (2004, 2005). Although it is now common practice to apply panel data unit
root and stationarity tests that take into account cross-section dependence, few
really test whether the individuals are cross-section dependent.
In this subsection we test the null hypothesis of non correlation against the
alternative hypothesis of correlation using the approach suggested in Pesaran
(2004) and Ng (2006). Besides, this framework allows us to gain some insight on
the kind of cross-section dependence in terms of how pervasive and strong is
the cross-section correlation. We can allow for the presence of the structural
breaks when testing the null hypothesis of non correlation among the individuals
in the panel. We will then estimate an autoregressive model to isolate cross­
section dependence from the autocorrelation that might be driving the
individual time series. In addition, the estimation of the autoregressive model
includes dummy variables to capture the level shifts that have been detected
using Bai and Perron (1998) in the previous section, which aims at isolating
cross-section dependence from both autocorrelation and structural breaks in
the individual time series.
Pesaran (2004) designs a test statistic based on the average of pair-wise
Pearson's correlation coefficients p̂ j , j = 1,2,K,n , n = N(N − 1) / 2 , of the residuals
— 34 — Instituto de Estudios Fiscales
obtained from an autoregressive (AR) model that includes dummy variables to
capture the structural breaks. The CD statistic of Pesaran (2004) is given by
CD =
2T
n
n
∑ p̂ j → N(0,1.)
j =1
This statistic tests the null hypothesis of cross-section independence against
the alternative of dependence.
The procedure proposed by Ng (2006) works as follows. First, we get rid of
the autocorrelation pattern in the individual time series through the estimation
of an AR model. This allows us to isolate the cross-section regression from serial
correlation. Taking the estimated residuals from the AR regression equations as
individual series, we compute the absolute value of Pearson's correlation
coefficients (p j = p̂ j ) for all possible pairs of individuals, j = 1,2,K,n , where
n = N(N − 1) / 2 , and sort them in ascending order. As a result, we obtain the
sequence of ordered statistics given by {p[1 : n], p[2 : n],K, p[n : n] }. Under the
null hypothesis that p j = 0 and assuming that individual time series are Normally
distributed, p j is half-normally distributed. Furthermore, let us define φ j as
(
), where Φ
denotes the cdf of the standard Normal distribution, so
that φ = (φ1,K, φn ) . Finally, let us define the spacings as ∆φ j = φ j − φ j−1 , j = 1,K,n .
Φ Tp[j
: n]
Second, Ng (2006) proposes splitting the sample of (ordered) spacings at
arbitrary ϑ ∈ (0,1) , so that we can define the group of small (S ) correlation
coefficients and the group of large (L ) correlation coefficients. The definition of
the partition is carried out by minimizing the sum of squared residuals
Q n (ϑ) =
[ϑn]
n
j=1
j=[ϑn]+1
2
2
∑ (∆φ j − ∆ S (ϑ)) + ∑ (∆φ j − ∆L (ϑ)) ,
where ∆ S (ϑ) and ∆ L (ϑ) denotes the mean of the spacings for each group
respectively. A consistent estimate of the break point is obtained as
ϑ̂ = argmin ϑ∈(0,1) Q n (ϑ) , where some trimming is required. Following Ng (2006)
the trimming is set at 0.10.
Once the sample has been split, we can proceed to test the null hypothesis
of non correlation in both sub-samples. Obviously, the rejection of the null
hypothesis for the small correlations sample will imply also rejection for the
large correlations sample as the statistics are sorted in ascending order.
Therefore, the null hypothesis can be tested for the small, large and the whole
sample using the Spacing Variance Ratio (SVR ) in Ng (2006), which under the
null hypothesis converges to the standard normal distribution.
The results in Tables 5 show that for ca i, t the null hypothesis of
independence is rejected for the whole sample of spacings, while it is not
— 35 — rejected for the L and S samples at the 5% level of significance. Since the
proportion of non significant correlations in the L and S group is similar, this
leads us to conclude that cross-section dependence is not pervasive. In this case,
the factor models suggested by Bai and Ng (2004) might not be a suitable
approximation to account for the cross-section dependence that appears in the
panel data set. Besides, Pesaran's (2004) CD statistic strongly rejects the null
hypothesis of independence.
As for the variable nfa i,t , we can see that the SVR statistic does not reject
the null hypothesis of cross-section independence at the 5% level of significance
neither of the whole, small and large sample of spacings. On the contrary,
Pesaran's (2004) CD statistic strongly rejects the null hypothesis of cross-section
independence. Although the two statistics lead to contradictory results, we
prefer to proceed in a conservative way and consider that the time series of nfa& i,t
in the panel data set are cross-section dependent – further note that the p-value
of the SVR test is 0.124, not far away from the 10% level of significance.
In all, the evidence that is obtained in this section indicates that cross-section
dependence has to be considered when computing the panel data statistics if
misleading conclusions are to be avoided.
5.1.4. Panel data tests with cross-section dependence and structural breaks
The specification estimated above permits the computation of two different
panel data stationarity statistics. First, we have applied the approach suggested
in Carrion-i-Silvestre et al. (2005) to test the null hypothesis of panel variance
stationarity allowing for multiple level shifts. Thus, note that the specification
given in (17) is one of the two models considered by these authors. The OLS
estimated residuals from (17) are used to obtain the individual KPSS statistics
computed in the previous sections, which in turn can be combined to define the
panel stationarity test statistic:
LM(λ ) = N −1
N
∑ η̂i (λ i ),
i =1
with η̂i (λ i ) defined in (18). Note that η̂i (λ i ) has been defined such that the long­
run variance is heterogeneous across individuals. However, it would be possible
to use a homogeneous estimate of the long run variance, i.e., ω̂ 2 = N −1 ∑Ni=1 ω̂i2 .
Using these elements we can define the panel data statistic
2
−1 N 2
2
Z(λ ) = N (LM(λ ) − ξ ))ς , where ξ = N −1 ∑N
i=1 ξ i and ς = N ∑i=1 ς i , with ξ i and ς i
being the individual mean and variance of ηi (λ i ) respectively. Note that these
two possibilities for the definition of the long-run variance estimate gives rise to
two different statistics, i.e., the Z(λ ) when the long-run variance homogeneity is
imposed and the Z(λ ) for heterogeneous long-run variance.
— 36 — Instituto de Estudios Fiscales
Under the null hypothesis of variance stationarity and assuming cross-section
independence, the Z(λ ) panel data statistics are shown to converge to the
standard normal distribution. However, this limiting result is not obtained when
individuals are cross-section dependent, as it is in our case. In this situation, we
can compute the bootstrap distribution of the Z(λ ) statistics to account for the
presence of a general form of cross-section dependence. The computation of
the bootstrap distribution follows the lines given in Maddala and Wu (1999). To
be specific, we have defined the (T × N) -matrix of the OLS estimated residuals
from (17) ê = (ê1,K,êN ) , and have resampled with replacement the rows of the
ê matrix so that the first matrix of resampled residuals ê ∗(1) is obtained, where
the superscript ∗ (1) indicates the first resampling. Conditional on the estimated
parameters and structural breaks, we have computed the bootstrap variables
∗(1)
y i,t
= α̂ i +
m̂i
∑ θ̂i,k DUi,k,t + ei,t∗(1),
k=1
for each i , where α̂ i and θ̂i, k are the OLS estimates of the parameters in (17).
∗(1)
This is repeated 20,000 times so that we define y i,t
,K, y ∗i,t(2,000 ) series for each
individual, which can be used to approximate the empirical distribution of the
Z(λ ) statistics. Table 1, panel B, presents the Z(λ )
statistics as well as the
bootstrap critical values for ca it . According to these statistics, the null hypothesis of I(0) cannot be rejected for the current account imbalance,
regardless of the assumption made about the long-run variance estimation.
Concerning the net foreign assets, the results are presented in Table 4, where we reach the same conclusion: the variable is stationary. Although we have already obtained that the variables in the two panels are
I(0) stationary processes, we have checked the robustness of our results
computing panel data unit root and stationarity tests that control for the
presence of cross-section dependence using approximate common factor
models proposed in Bai and Ng (2004), Harris et al. (2005) and Bai and Carrion­
i-Silvestre (2009). The common factors approach decomposes the observable
variables as follows
y i, t = α i +
mi
∑ θi, k DUi, k, t + Ft′ πi + ξi, t ,
(19)
k =1
t = 1,K,T , i = 1, K,N , where Ft is a (r × 1) -vector that accounts for the common
factors that are present in the panel, and ξi, t is the idiosyncratic disturbance
term, which is assumed to be cross-section independent. Note that the
specification given by (19) is similar to the one in (17), where the disturbance
term e i,t in (17) has been expressed as e i, t = Ft′ π i + ξ i, t giving rise to the
specification in (19). The unobserved common factors (Ft ) and idiosyncratic
— 37 — disturbance terms (ξi, t ) are estimated using principal components on the first
difference model. The estimation of the number of common factors is obtained
using the panel BIC information criterion in Bai and Ng (2002), with a maximum
of six common factors. Tables 1 and 4, respectively, report the results of
applying this method. For both the current account and the net foreign assets
the ADF statistic computed from the idiosyncratic disturbance terms rejects the
null hypothesis of unit root, while the procedure detects at least one non­
stationary common factor (one in cai, t , and six in nfa i,t ) where r1 denotes the
number of non-stationary common factors so that r = r0 + r1 , and r0 is the
number of stationary common factors.
This set-up allows us to compute two panel data test statistics that consider
the presence of multiple structural breaks. First, we have the panel data
stationarity test statistic in Harris et al. (2005), which is given by
ˆ = T −1/ 2 ∑T
ˆ + cˆ ) / ω
ˆ {aˆ k,t }, being C
S Fk = (C
k
k
t =k +1 â k,t the autocovariance of order k ,
+r̂
â k,t = ∑N
i=1 ẑ i,t ẑ i,t −k . We define ẑ i, t as the i th element of the (N + r̂ ) × 1 vector
(F̂1, t ,K,F̂r̂, t ,ξ̂1, t ,K,ξ̂N, t )′
that contains the estimated common factors (F̂) and the
idiosyncratic disturbance (ξˆ i ) , with ĉ = (T − k )−1/ 2 ∑Ni=1ĉ i , being ĉ i a correction
term defined in Harris et al. (2005) and, ω̂2 {a t } is a consistent estimate of the
long-run variance of {a t } . Under the null hypothesis of joint variance stationarity
of the common and idiosyncratic components the statistic SFk → d N(0,1) . We
follow Harris et al. (2005) and use k = (3T )1/ 2 . The value of S Fk = 2.546 statistic
with p-value of 0.005 leads to the rejection of the null hypothesis of I(0), which
contradicts the previous results that have been found using the Z (λ ) statistics.
However, it should be borne in mind that the Ng's statistic has revealed that
cross-section dependence is not pervasive, so that the use of the common
factor model might not be correct.
Second, we can compute the panel data unit root test in Bai and Carrion-iSilvestre (2009), which has been shown to be robust to the presence of multiple
structural breaks affecting the level. These authors propose the computation of
the panel data version of the modified Sargan-Bhargava (MSB) statistics using the
estimated idiosyncratic disturbance term (ξ̂i ), with up to three different ways to
pool the individual information. In this case and using the notation in Bai and
Carrion-i-Silvestre (2009), we have the Z → d N(0,1) , Pm → d N(0,1) and P → d χ 22N
panel data unit root test statistics. Panel B of 1 reports the results for the
variable ca i, t , showing that the null hypothesis of I(1) can be rejected using the
Pm and P statistics at the 5% level of significance, although it is not rejected
when using the Z test. The finite sample analysis in Bai and Carrion-i-Silvestre
(2009) shows that the Pm and P statistics are the ones with better finite sample
[
]
— 38 — Instituto de Estudios Fiscales
performance compared to the Z test, as the Z test suffers from mild size
distortions problems (underrejection) while the Pm and P statistics have the
correct size. Thus, we rely on the conclusions of the Pm and P statistics, that
the null hypothesis of panel unit root for ca i, t is rejected.
We have conducted the same analysis for the nfa variable, whose own
results are reported in Panel B of Table 4. In this case we present the statistics
for different number of common factors, given that the use of the panel BIC
always estimates the number of common factors that equals the maximum that
is allowed. First, the SFk statistic does not reject the null hypothesis of I(0)
regardless of the number of common factors that is specified. Second, the
conclusions drawn from the panel unit root tests depend on the number of
common factors that are considered. In general, we can see that when we
impose either one, three, four or five common factors, the null hypothesis of unit
root is rejected at least at the 10% level of significance. The converse is obtained
when the number of common factors is fixed at two or six. Taken together, the
results that have been obtained indicate that nfa i,t is I(0) stationary.
To sum up, our results show that there is evidence that both the current
account and the net foreign assets variables can be characterized as I(0) stationary
processes once structural breaks and cross-section dependence are allowed for.
5.2. Testing for the unified approach
5.2.1. Bai-Perron estimation results
Following the discussion in the section devoted to the theoretical models, we
now proceed to test for what we have called the “unified approach”. For this
purpose we have estimated the following model:
CA t
NFA t
= µj + βj
+ ut ;
GDPt
GDPt
t = Tj−1 + 1,...,Tj
j = 1,...,m + 1, where T0 = 0 and Tm+ q = T. Given that the variables that define this
model have been characterized as I(0) stochastic processes, we can apply the
Bai and Perron (1998) procedure to estimate the number and position of the
structural breaks. The reason for the choice of this method is that, due to the
presence of an important number of structural changes in the unit root analysis,
we expect to find also breaks in the relationship linking ca t and nfa t . In fact,
only some of the changes cancel-out and the majority remain. In Table 6 we
present the results for all the countries in our sample. We show in the first four
columns the dates where we have found the structural changes. In the following
columns we offer the results for the mean and the slope parameters in the
— 39 — different regimes. There are five cases where no structural change was found:
France, Greece, New Zealand, Portugal and Spain. Moreover, the relationship is
not significant in France and in New Zealand. As all of them had breaks
individually, we assume that they cancel-out.
A second group of countries has only one structural break: Australia (1976),
Japan (1987) and the Netherlands (1993). Probably the changes in regime are
related to exchange rate movements. In particular, the Dutch one is found in
the 1993 EMS crisis. Two structural breaks appear in Finland (1988, 1994),
Ireland (1982, 1996), Italy (1992, 1998), Sweden (1993, 2001) and the UK
(1981, 1986). Whereas the EMU members changes are linked to the transition
towards the new regime (Italy, Finland and Ireland), in the UK the oil prices are
the main determinants of the changes in its external position. The rest of the
countries have three structural breaks except for the US, which has four.
As we have already described in the previous sections, the relation between
the two measures of the external position of the countries should be positive.
This is what we find: the structural changes are mainly capturing the activation
of the adjustment mechanisms in the face of external imbalances. However, the
parameters in the different regimes maintain the positive link between the two
series. Take, for example, the case of Denmark in Table 6: all the β̂ are
significant and positive. This is also true for the majority of the countries, no
matter the number of breaks or its EMU membership.
This is the first step of this part of the analysis, where we have estimated the
reduced-form equation for all the countries in the sample. In the next
subsection we will study the causality and the dynamics of the adjustment for
two of the countries: the US and Spain. Note that the representation that we
will use only concerns the dynamics or short-run analysis. We have restricted
the last issue of the study to just two countries that are those (together with
Portugal) in our sample with the larger current account deficits from 2000
onwards and where the current account adjustment seems to have been
sluggish.
5.2.2. VAR and impulse-response results: the case of US and Spain
We have now investigated the Granger causality between ca and nfa
variables through the estimation of a VAR(p) model specification, allowing for
structural breaks that affect both the constant and the coefficients of the lagged
variables. Our analysis relies on the previous results where the two variables
under investigation have been characterized as I(0) stationary processes. In this
case we can apply the procedure in Qu and Perron (2007) to estimate a system
of equations that allow for the presence of common structural breaks affecting
the different equations.
— 40 — Instituto de Estudios Fiscales
Let us define the vector of variables y t = (ca t ,nfa t )′ . The model that has been
estimated using Qu and Perron (2007) approach is given by:
y t = µ j + β j,1y t −1 + β j,2 y t −2 + L + β j,p y t −p + u t ;
t = Tj−1 + 1,K,Tj
where β j,k denotes the (2 × 2) matrix of parameters that is associated to the k -th
lag in the j -th regime, k = 1,K,p , j = 1,K,m . Given the limitation imposed by the
small number of observations that we dispose of, we have restricted both the
number of structural breaks and the maximum lag order for the VAR model up
to two. Further, the covariance matrix of the disturbance terms is assumed not
to change across regimes. The order of the VAR model is chosen using the
Bayesian information criterion (BIC).
Spain
We should first note that in the Bai-Perron procedure applied to the
reduced-form model (or structural representation) no breaks were found in the
case of Spain. The dynamics differ: two structural changes (1985 and 1995)
were found. This finding does not contradict the above-mentioned stability. It
only calls for different types of adjustment in the three regimes. The first one,
previous to EU accession; a second one between 1986 and the application of
the Maastricht criteria; the third one after 1996 (roughly EMU membership).
The estimates for the Spanish case using the whole period that goes from
1972 to 2006 are the following ones:
⎛ − 2.749 ⎞ ⎡0.592
⎜⎜
⎟⎟ + ⎢
⎝ − 1.767 ⎠ ⎣0.702
⎛ − 3.438 ⎞ ⎡ 0.661
⎜⎜
⎟⎟ + ⎢
⎝ − 3.283 ⎠ ⎣0.402
⎛ ca ⎞ ⎛ 0.562 ⎞ ⎡0.730
⎜⎜ t ⎟⎟ = ⎜⎜
⎟⎟ + ⎢
⎝ nfa t ⎠ ⎝ − 3.081⎠ ⎣2.106
⎛ ca t ⎞
⎜⎜
⎟⎟ =
nfa
t
⎠
⎝
⎛ ca t ⎞
⎟⎟ =
⎜⎜
⎝ nfa t ⎠
− 0.265 ⎤ ⎛ ca t −1 ⎞
⎜
⎟⎟ + û t
0.767 ⎦⎥ ⎝⎜ nfa t −1 ⎠
t = 1972,K,1985
− 0.188 ⎤ ⎛ ca t −1 ⎞
⎟⎟ + uˆ t
⎜
0.776 ⎦⎥ ⎝⎜ nfa t −1 ⎠
t = 1986,K,1995
0.075 ⎤ ⎛ ca t −1 ⎞
⎜
⎟ + uˆ t
0.824⎦⎥ ⎝⎜ nfa t −1 ⎠⎟
t = 1996,K,2006
These estimates satisfy the stationarity condition for the first and second
regime, but not for the third regime. Although this might seem to contradict our
previous results, we should bear in mind that the analysis is constrained to
consider at most two structural breaks. Thus, the procedure that we have
applied does not allow to include a third structural break in the model because
of the small number of observations that we have, a situation that might be
behind the non-stationarity of the VAR model in the third regime, specially, if
we take into account that the euro was launched in 1999.
In order to evaluate this suspicion, we have carried out the estimation of the
VAR model using the observations up to 1999. The estimates of the model for
this subperiod are:
— 41 — ⎛ ca ⎞
⎜⎜ t ⎟⎟ =
⎝ nfa t ⎠
⎛ ca t ⎞
⎜⎜
⎟⎟ =
⎝ nfa t ⎠
⎛ − 2.589 ⎞ ⎡0.569 − 0.229 ⎤ ⎛ ca t −1 ⎞
⎟⎟ + û t t = 1972,K,1982
⎜⎜
⎟⎟ + ⎢
⎥ ⎜⎜
⎝ − n1.610 ⎠ ⎣0.797 0.769 ⎦ ⎝ nfa t −1 ⎠
⎛ − 5.087 ⎞ ⎡ 0.911 − 0.371⎤ ⎛ ca t −1 ⎞
⎜⎜
⎟⎟ + ⎢
⎥ ⎜⎜
⎟⎟ + uˆ t t = 1983,K,1993
⎝ − 3.739 ⎠ ⎣0.559 0.698 ⎦ ⎝ nfa t −1 ⎠
⎛ ca ⎞
⎜⎜ t ⎟⎟ =
⎝ nfa t ⎠
⎛ − 3.037 ⎞ ⎡1.360 − 0.163 ⎤ ⎛ ca t −1 ⎞
⎜⎜ − 26.844 ⎟⎟ + ⎢ 5.051 − 0.623 ⎥ ⎜⎜ nfa ⎟⎟ + uˆ t
⎝
⎦⎝
⎠ ⎣
t −1 ⎠
t = 1994,K,1999
Now the estimated VAR models for each subperiod satisfy the stationarity
conditions. We can use these estimates to compute impulse response functions
(IRF) to evaluate the effects of the shocks of each equation.
Figures 11 to 13 present the IRF that are based on the Cholesky's orthogonal
decomposition of the covariance matrix of the disturbance terms. As can be
seen, the same qualitative picture is obtained for the first two regimes. Thus, we
can see that the shocks that affect the current account, no matter from which
equation they come, cause a decrease in the current account in the long-run.
The opposite is observed for the effects on nfa t , for which the shocks cause an
increase in the variable nfa t in the long-run. However, the main difference
between the two regimes comes from the magnitude of these effects: while the
magnitude of the effects on the current account has increased, whereas the one
on nfa t has decreased.
This situation changes for the third regime, where a shock (of magnitude equal
to one standard deviation) coming from the disturbance that affects the CA
equation produces a positive effect both on ca t and nfa t in the long-run –the
total multiplier for ca t equals 6.915 and for nfa t is 22.274. On the other hand,
when the shock comes from the disturbance that affects the nfa t equation it
causes a negative effect on both ca t and nfa t variables in the long-run– the total
multipliers are -0.925 and -2.050, respectively.
US
The estimated model for the US using the statistical information for the
whole period gives the following results:
⎛ ca ⎞
⎜⎜ t ⎟⎟ =
⎝ nfa t ⎠
⎛ ca ⎞
⎜⎜ t ⎟⎟ =
⎝ nfa t ⎠
⎛ 1.348 ⎞ ⎡0.414 − 0.224 ⎤ ⎛ ca t −1 ⎞
⎜⎜
⎟⎟ + ⎢
⎥ ⎜⎜ nfa ⎟⎟ + û t t = 1972,K,1982
3.130
0.053
0.545
⎝
⎠ ⎣
⎦⎝
t −1 ⎠
⎛ − 1.256 ⎞ ⎡0.577 − 0.115 ⎤ ⎛ ca t −1 ⎞
⎟⎟ + û t t = 1983,K,1995
⎜⎜
⎟⎟ + ⎢
⎥ ⎜⎜
⎝ − 2.024 ⎠ ⎣0.070 0.694 ⎦ ⎝ nfa t −1 ⎠
⎛ ca ⎞
⎜⎜ t ⎟⎟ =
⎝ nfa t ⎠
⎛ − 0.596 ⎞ ⎡0.854 0.028 ⎤ ⎛ ca t −1 ⎞
⎜⎜
⎟⎟ + ⎢
⎥ ⎜⎜
⎟⎟ + û t
⎝ − 4.807 ⎠ ⎣0.927 0.512⎦ ⎝ nfa t −1 ⎠
— 42 — t = 1996,K,2006
Instituto de Estudios Fiscales
It can be verified that these estimates satisfies the stationarity conditions for
each regime, whose IRF are reported in Figures 14 to 16. The structural breaks,
although just two (note that the structural model in the previous section had
four breaks) capture the same type of discontinuities than above. The changes
occur in 1982 and 1995. As can be seen, the same qualitative conclusions are
obtained for the first and second regimes. Thus, the shocks that come from the
disturbance of the ca t equation affect positively the ca t in the long-run,
whereas the effect is negative for the nfa t . Besides, the shocks coming from the
disturbance of the nfa t equation affects positively nfa t itself in the long-run,
whereas the effect is negative on the current account. An interesting feature is
that the effects of the shocks seem to favor the net foreign asset position of the
US when going from the first to the second regime: (i) a shock coming from the
disturbance of the ca t equation reduces the nfa t by -0.109 in the first regime
and by -0.048 in the second regime, while (ii) a shock coming from the
disturbance of the nfa t equation increases the nfa t by 2.839 in the first regime
and by 4.154 in the second regime.
As for the third regime, the results indicate that the effect of the shocks is
always positive for all variables.
6. CONCLUSIONS
In this paper we have empirically revisited the debate of the external
accounts sustainability in the OECD countries for the period 1970-2006. We try
to reconcile the main theoretical approaches and formulate hypotheses that can
be tested in a panel framework. Current account imbalances have steadily
increased in rich countries over the last 20 years and there appears a widely
shared worry that these deficits are too large, and government intervention is
required. Using the concept of sustainability as the ability to meet the long run
intertemporal budget constraint, we test for stationarity in the current account
and the stock of net foreign assets of the OECD countries. In addition, we call
for a unified approach were we relate both the flow and the stock approaches
in line with Gourinchas and Rey (2007).
Concerning the empirical methodology, we argue that there are several
reasons to believe that these variables may suffer from discontinuities. Previous
evidence has revealed that there might be some events that affect the current
account and the external debt in a permanent way. If this is the case, it is well
known that non accounting for structural breaks biases both unit root and
stationarity tests towards concluding in favor of non-stationarity in variance.
Moreover, the independence assumption imposed in the so-called first
— 43 — generation panel data statistics has been widely criticized in the recent
literature, since it has been shown that non accounting for cross-section
dependence amongst the individuals might bias the statistical inference in favor
of I(0) stationarity.
In this research we aim at filling the gap in the literature on external
sustainability in several respects. First, we improve previous empirical work on
the intertemporal model by testing for the stationarity of the current account
and the net foreign assets stock by applying panel tests. Second, we allow for
multiple structural breaks and cross-section dependence. Finally, we relate the
identification of the structural changes with the literature on current account
reversals, trying to assess how the countries regain solvency through adjustment
processes.
Concerning the classical flow approach, based on the current account, our
results point at just two cases of strict sustainability: only two countries have not
experienced structural changes during the analyzed period, namely, Italy and
New Zealand, showing stationarity, and therefore, external sustainability. The
rest of the countries have experienced up to four breaks in their current
account for the period considered. These discontinuities correspond to major
institutional changes or policy measures that have induced a series of breaks in
the path followed by the variables. Focusing on the individual statistics, we can
see that the null hypothesis of I(0) cannot be rejected at the 5% level of
significance for fifteen out of twenty countries – the exceptions are Ireland,
Japan, Netherlands, Portugal, and Sweden. In general, the individual country
results point to the fact that policy measures or, otherwise, abrupt
readjustments, are still needed to keep the sustainability of the current
accounts. This evidence would be against a smooth self regulating capacity of
the markets, and therefore, against laissez-faire, the so-called Lawson doctrine.
However, the increasing financial integration process among the OECD
countries may be relaxing the external constraint. In fact, the evidence obtained
indicates that cross-section dependence has to be considered when computing
the panel data statistics if misleading conclusions are to be avoided. Finally, our
results show that there is evidence of the current account being an I(0)
stationary process once structural breaks and cross-section dependence are
allowed for.
The stock approach has been tested using the same empirical methodology.
Again the results show that, even though the whole panel turns out to be
stationary once we account for structural changes and dependence, only one
country, namely Germany, does not suffer from structural changes.
Once the two variables are found to be stationary, we analyze whether there
is a relationship linking them, their sign, stability and dynamics. We first assess
the presence of structural changes in the relation using the Bai-Perron
— 44 — Instituto de Estudios Fiscales
procedure and estimate the reduced-form parameters for the subperiods
defined by the structural changes. For the majority of the countries and time­
periods the parameters are positive, smaller than one and significant, as
expected. Concerning the dynamics, we have applied a recent method
proposed by Qu and Perron (2007), which consist of the estimation of a VAR
model also allowing for structural changes. We restrict this analysis to the cases
of Spain and the US, which are those where the external imbalances are more
severe. The results point to stationary adjustment for the majority of the
sample, the exception being Spain after 1996. This implies that solvency is
recovered after major shocks that affect the countries' external accounts.
However, the high degree of financial globalization may have increased the
persistence of the disequilibria for some peripheral EMU members, as in the
case of Spain.
— 45 — Instituto de Estudios Fiscales
APPENDIX
Figure 1 TRADE OVER GDP 80
T RAEMU
T RAUK
T RAUS
70
T RAJ AP
60
50
40
30
20
10
0
1 960 19 63 19 66 196 9 1 972 19 7 5 197 8 1 981 1 984 19 87 199 0 1 9 93 19 9 6 1 99 9 2 0 02 200 5
Figure 2 FINANCIAL INTEGRATION
Fina ncial integration= asse ts + liabilities o ver GDP
2500
900
FI PORT UGA L
FI ITA L Y
FI GRE E C E
FI SPA I N
FI IRE L A ND
FIU S
FIC A NA DA
FIU K
FIN E WZE
FIA U ST RA LI A
800
2000
700
600
1500
500
400
1000
300
200
500
100
0
0
1970
1975
1980
1985
1990
1995
2000
2005
1970
1975
1980
1985
1990
1995
2000
2005
2000
2005
Figure 3 FINANCIAL INTEGRATION
Financi al in tegration= assets + liab ilitie s over GDP
60 0
1000
FI SW E D EN
FI NOR W AY
FI DE N MA R K
FI FIN L AN D
FI AU S TR IA
50 0
FI FR AN C E
FI GE R
FI NE TH
FI BE LG IU M
FI JA PA N
800
40 0
600
30 0
400
20 0
200
10 0
0
0
1970
197 5
1980
19 85
1990
1 995
2000
2005
1 970
— 47 — 1 975
1980
1985
1990
1995
Figure 4 CURRENT ACCOUNT
CA ove r GD P
5.0
4
CA POR
CA IT
CA GR E
CA ES P
CA IR E
2.5
CA US
CA CAN
CA UK
CA NZ
C A A US L
2
0.0
0
-2.5
-2
-5.0
-4
-7.5
-6
-10.0
-8
-12.5
-10
-15.0
-12
-17.5
-14
1970
1975
1980
1985
1990
1995
2000
2005
1970
1975
1980
1985
1990
1995
2000
2005
1980
1985
1990
1995
2000
2005
1980
1985
1990
1995
2000
2005
Figure 5 CURRENT ACCOUNT
CA over GDP
20
10.0
CA SW E
CA NOR
CA DK
CA FIN
CA AU S
15
CAFR
CAGE R
CANE T
CAB E L
CAJ A P
7.5
5.0
10
2.5
5
0.0
0
-2.5
-5
-5.0
-10
-7.5
-10.0
-15
1970
1975
1980
1985
1990
1995
2000
2005
1970
1975
Figure 6 NET FOREIGN ASSETS
NFA over GD P
50
25
NFAP OR
NFAI T
NFAGR E
N FAE S P
NFAI R E
25
NFA US
NFA CA N
NFA UK
NFA NZ
NFA AU S L
0
0
-25
-25
-50
-50
-75
-75
-100
-100
-125
1970
1975
1980
1985
1990
1995
2000
2005
1970
— 48 — 1975
Instituto de Estudios Fiscales
Figure 7 NET FOREIGN ASSETS
NFA over GDP
100
75
NFAS W E
NFAN OR
NFAD K
NFAFI N
NFAA U S
50
NFAFR
NFAGE R
NFAN E T
NFAB E L
NFAJ A P
50
0
25
-50
0
-100
-25
-150
-200
-50
1970
1975
1980
1985
1990
1995
2000
2005
1970
1975
1980
1985
1990
1995
2000
2005
Table 1
RESULTS FOR THE MODEL WITH MULTIPLE BREAKS AFFECTING THE MEAN
Panel A: Individual information
Tests
mi
i
Tb,1
Australia
Austria
Belgium
Canada
Denmark
Finland
0.122
0.037
0.031
0.057
0.073
0.067
1
2
4
1
1
1
1980
1975
1974
1998
1989
1994
France
Germany
Greece
Ireland
0.041
0.034
0.096
1
3
1
3
1992
1984
1998
1984
Italy
Japan
Netherlands
New Zealand
0.181 ∗∗
0.035
0
2
1982
0.256 ∗∗
0.109
1
1992
0.053
1979
1984
1999
0.221 ∗∗
i
Tb,
2
1981
1984
0.160 ∗∗
2
2
Spain
0.269 ∗
1
1992
1990
2001
1991
1998
2001
0
Norway
Portugal
i
Tb,
3
1999
1995
i
Tb,
4
2000
Critical values
10%
5%
0.197
0.179
0.060
0.230
0.156
0.186
0.256
0.232
0.068
0.302
0.190
0.237
0.169
0.091
0.231
0.085
0.212
0.110
0.299
0.102
0.354
0.128
0.463
0.158
0.168
0.209
0.350
0.453
0.127
0.100
0.161
0.118
0.244
0.319
(Follow)
— 49 —
(Continuation)
Panel A: Individual information
Tests
mi
i
Tb,1
i
Tb,
2
Sweden
0.428 ∗∗
2
1994
United Kingdom
United States
0.086
0.042
1
2
1986
1982
i
Tb,
3
i
Tb,
4
Critical values
10%
5%
2001
0.180
0.233
1999
0.158
0.114
0.191
0.138
Panel B: Panel data based unit root and stationarity test statistics
Bootstrap distribution
Test
90%
95%
Z(λ) (Homog)
-0.521
4.620
5.853
Z(λ ) (Heterog)
-1.904
4.882
5.617
Test
p-value
Num. of
factors
SFk
12.546
0.005
1
Z
Pm
-0.735
11.921
0.231
0.027
5
5
P
57.178
0.038
5
Table 2 MAIN EVENTS AND BREAKS FOUND IN THE DATA
EMU countries
Main events
CA Surplus
Beginning 70s
(Bretton Woods ends)
AUS (75)
First oil shock
BEL (74)
Both
Non-EMU countries
CA
Deficit
CA surplus
CA
deficit
Beginning 80s
Second oil shock
AUS (81)
NOR(79)
IRE(84)
Mid-80s
lower oil prices
BEL(84), GER(84)
Beginning-mid 90s
BEL(92)
German unification,
K mov, EMS crises
GER(90), FIN(94) FR(92) POR(95) DK(89), SWE(94)
End 90s, beginning 2000
Asian Crisis, EMU
AUST(80)
POR(84)
IRE(91)
GRE(98) CAN(98), NOR(99)
BEL(00), GER(01) IRE(98) SPA(99)
— 50 — SWE(01), JAP(01) US(99)
Instituto de Estudios Fiscales
Table 3
MAIN EVENTS AND BREAKS FOUND IN THE DATA
Main events
Non-EMU
countries
EMU countries
Negative NFA
Positive NFA
Negative NFA
Positive NFA
Beginning 70s
(Bretton Woods
ends)
NZ (76 ↓ ),
CAN(77 ↓ )
First oil shock
Beginning 80s
AUS(76)
BEL(80), FR(82)
Second oil shock
Mid-80s
GRE(80)
SWE(80 ↓ )
NZ(81), DK(83)
IRE(80), SP(80)
UK(80 ↑ )
AUST(78)
JAP(84), US(84)
ITA(84 ↓ )
lower oil prices
Beginning-mid
90s
DK(77)
IRE(88)
BEL(90,96)
AUST(85)
ESP(92),AUS(96)
JAP(90,95)
NOR(90 ↑ ),DK(91),
NZ(91)
UK(89,95 ↓ )
SWE(92), US(94 ↓ )
POR(00),ITA(01)
SWE(97 ↑ )
DK(97), CAN(98 ↑ ),
NZ(99)
GRE(01),SPA(01)
JAP(00),
NOR(01)
AUST(01),
US(99 ↓ )
German unification, NET(93 ↓ ),FR(95),IRE(96 ↑ ),FIN(96 ↓ )
K mov, EMS crises
End 90s,
beginning 2000
FIN(01 ↑ )
Asian Crisis, EMU
Table 4 RESULTS FOR NFA VARIABLE WITH MULTIPLE BREAKS AFFECTING THE MEAN
Panel A: Individual information
Tests
mi
i
Tb,1
i
Tb,
2
i
Tb,
3
2001
i
Tb,
4
Critical values
10%
5%
Australia
0.098
3
1978
1985
0.099 0.122
Austria
0.214 ∗∗
2
1976
1996
Belgium
0.129 ∗∗
3
1980
1990
Canada
0.105
2
1977
1998
Denmark
0.046
4
1977
1983
Finland
0.097
2
1996
2001
0.196 0.255
France
0.048
2
1982
1995
0.098 0.113
0.142 0.179
1996
0.075 0.085
0.149 0.188
1991
1998
0.056 0.063
(Follow)
— 51 —
(Continuation)
Panel A: Individual information
Tests
mi
i
Tb,1
i
Tb,
2
i
Tb,
3
i
Tb,
4
Critical values
10%
5%
Germany
0.092
0
0.356 0.459
Greece
0.054
2
1980
2001
Ireland
0.042
3
1980
1988
Italy
0.239 ∗∗
2
1984
2001
Japan
0.517 ∗∗
4
1984
1990
Netherlands
0.076
1
1993
New Zealand
0.052
4
1976
1981
Norway
0.025
2
1990
2001
Portugal
0.033
1
2000
Spain
0.439 ∗∗
3
1980
1992
2001
0.079 0.092
Sweden
0.201 ∗∗
3
1980
1992
1997
0.079 0.092
United Kingdom
0.123 ∗∗
3
1980
1989
1995
0.075 0.086
United States
0.123 ∗∗
3
1984
1994
1999
0.084 0.099
0.151 0.193
1996
0.071 0.081
0.123 0.149
1995
2000
0.073 0.088
0.169 0.209
1991
1999
0.061 0.071
0.136 0.169
0.252 0.330
Panel B: Panel data based unit root and stationarity test statistics
Bootstrap distribution
Test
90%
95%
Z(λ)
(Homog)
-1.349 13.442 15.204
Z(λ )
(Heterog)
-5.898 17.010 17.824
Number of factors
SFk
p-value
Z
p-value
Pm
p-value
P
p-value
1
-0.824 0.795 -1.300 0.097 4.194 0.000 77.512 0.000
2
-0.489 0.668 -1.271 0.102 0.664 0.253 45.942 0.240
3
-0.446 0.672 -2.092 0.018 2.431 0.008 61.747 0.015
4
-0.279 0.610 -1.591 0.056 1.645 0.050 54.712 0.061
5
-0.761 0.223 -1.229 0.110 2.569 0.005 62.976 0.012
6
-0.584 0.280 -1.049 0.147 0.533 0.297 44.766 0.279
— 52 — Instituto de Estudios Fiscales
Table 5 SPACING VARIANCE RATIO AND CD STATISTICS FOR THE CA AND NFA VARIABLES.
DETERMINISTIC FUNCTION GIVEN BY A CONSTANT TERM WITH LEVEL SHIFTS
Ng’s spacing test
Whole sample
ca
nfa
Pesaran’s CD test
Small group
Large group
svr (η)
p-val
svr (η)
p-val
η̂
svr (η)
p-val
Test
p-val
2.355
1.158
0.009
0.124
-0.674
-0.490
0.250
0.688
120
194
-0.859
-0.659
0.195
0.745
6.825
3.907
0.000
0.000
Figure 8 CURRENT ACCOUNT OVER GDP AND ESTIMATED BREAK POINTS
(Follow)
— 53 — (Continuation)
(Follow)
— 54 — Instituto de Estudios Fiscales
(Continuation)
— 55 — Figure 9 NET FOREIGN ASSETS OVER GDP AND ESTIMATED BREAK POINTS
(Follow)
— 56 —
Instituto de Estudios Fiscales
(Continuation)
(Follow)
— 57 — (Continuation)
— 58 — 1994
1988
Finland
Ireland
Greece
Germany
1982
1996
2001
1993
1986
1976
Denmark
1990
1993
1987
1981
Canada
1985
1999
1992
1984
Belgium
France
2001
1993
1981
Austria
Tˆb, 3
1976
Tˆb, 2
Australia
Tˆ b,1
Tˆb, 4
β̂ 1
µˆ 2
β̂ 2
-5.107 -0.307 -2.559 0.034
(-2.919) (-2.507) (-4.183) (2.660)
-0.722 0.274 0.461 0.044
(-3.550) (11.332) (1.279) (1.156)
-1.990 -0.129 2.337 0.021
(-9.232) (-5.842) (7.096) (0.412)
-3.500 -0.008 24.978 0.689
(-4.837) (-0.369) (11.006) (11.632)
11.282 0.677 -2.303 0.032
(6.711) (7.985) (-4.549) (2.447)
-6.491 -0.255 -10.324 -0.180
(-1.348) (-0.964) (-3.467) (-2.449)
0.215 0.059
(0.655) (1.124)
-3.050 0.588 4.044 -0.024
(-3.982) (4.923) (3.384) (-0.328)
-1.894 0.072
(-3.834) (4.310)
-0.373 0.174 5.831 0.117
(-0.886) (16.478) (14.181) (16.715)
µˆ 1
β̂ 3
µˆ 4
β̂ 4
-0.558 0.040
(-2.970) (4.795)
-0.835 -0.018 1.812 0.144
(-2.666) (-0.492) (3.606) (4.435)
-1.740 0.030 -5.626 -0.306
(-3.362) (1.083) (-4.945) (-5.891)
4.841 0.019 1.205 0.068
(5.766) (0.430) (1.124) (2.780)
0.196 0.096 3.137 0.117
(0.082) (1.573) (22.869 (20.314)
12.410 0.283 3.685 0.105
(11.703) (10.786) (17.772) (9.680)
5.845 -0.011
(6.004) (-0.875)
µˆ 3
Table 6 ESTIMATION OF THE CA AND NFA RELATIONSHIP
µˆ 5
(Follow)
β̂ 5
1993
Netherlands
1993
1981
1976
Sweden
UK
US
Spain
Portugal
Norway
1982
1986
2001
1985
1987
Japan
1977
1998
1992
Italy
New Zealand
Tˆb, 2
Tˆb,1
(Continuation)
1988
1999
Tˆb, 3
µˆ 1
β̂ 1
µˆ 2
β̂ 2
µˆ 3
β̂ 3
µˆ 4
β̂ 4
µˆ 5
β̂ 5
1.558 0.192
-0.211 0.144 4.841 0.347
(-1.513) (6.400) (6.572) (4.031) (3.134) (5.195)
-0.710 0.461 1.740 0.040
(-3.294) (11.543) (6.140) (3.608)
-1.901 0.280 6.675 0.116
(-1.001) (2.463) (9.895) (3.111)
-4.791 0.012
(-3.122) (0.520)
2.228 0.214 14.232 0.003
2.325 0.375 8.802 0.259
(1.570) (6.359) (5.771) (5.149) (5.086) (6.474) (7.524) (0.081)
-1.188 0.092
(-0.844) (2.656)
0.065 0.119
(0.132) (5.482)
5.161 -0.126
-0.607 0.049 2.581 0.025
(-1.211) (1.553) (1.427) (0.438) (8.662) (-2.410)
-1.963 0.425 3.510 -0.147 -2.215 -0.034
(-3.610) (4.038) (1.619) (-1.308) (-8.734) (-1.303)
-0.973 0.234 -3.884 0.511 -2.740 0.211 -0.183 0.212 -12.677 -0.381
2001
(-2.725) (3.703) (-4.615) (4.324) (-30.940) (8.590) (-1.410) (15.374) (-11.193) (-6.456)
Tˆb, 4
Instituto de Estudios Fiscales
Figure 10
CA AND NFA VARIABLES
(Follow)
— 61 — (Continuation)
(Follow)
— 62 —
Instituto de Estudios Fiscales
(Continuation)
Figure 11
SPAIN. ACCUMULATED IMPULSE RESPONSES FOR
THE FIRST REGIME (1972-1982) — 63 — Figure 12
SPAIN. ACCUMULATED IMPULSE RESPONSES FOR THE
SECOND REGIME (1983-1993) Figure 13
SPAIN. ACCUMULATED IMPULSE RESPONSES FOR THE
THIRD REGIME (1994-1999) — 64 — Instituto de Estudios Fiscales
Figure 14
US. ACCUMULATED IMPULSE RESPONSES FOR THE .RST REGIME (1972-1982) Figure 15
US. ACCUMULATED IMPULSE RESPONSES FOR THE
SECOND REGIME (1983-1995) — 65 — Figure 16
US. ACCUMULATED IMPULSE RESPONSES FOR THE
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— 72 — SÍNTESIS
PRINCIPALES IMPLICACIONES DE POLÍTICA ECONÓMICA
En este trabajo se ha realizado una revisión empírica del debate sobre la solvencia y
sostenibilidad externa de los países de la OCDE durante el período 1970-2006. Los
desequilibrios por cuenta corriente se han incrementado de forma constante durante
los últimos veinte años, existiendo una preocupación creciente sobre el tamaño que
han alcanzado y la necesidad de tomar medidas de ajuste más allá de los mecanismos
de mercado.
Con este fin se contrasta la hipótesis de solvencia, entendida como la capacidad de
un país para cumplir su restricción presupuestaria intertemporal, tanto desde el punto
de vista del saldo acumulado de deuda exterior (enfoque stock) como por los flujos de
la cuenta corriente en términos de valor presente (enfoque flujos). Asimismo, se
analiza la relación entre las dos variables dentro del enfoque unificado de Gourinchas y
Rey (2007).
Desde un punto de vista metodológico, se contrasta la hipótesis de estacionariedad
de las variables implicadas en un contexto de datos de panel no estacionarios,
realizando un conjunto de mejoras respecto a los trabajos empíricos precedentes. En
primer lugar, se amplia la información mediante el uso de datos de panel, lo que añade
información cross-section a la contenida en la dimensión temporal de los datos. En
segundo, se considera la existencia de cambios estructurales. Este punto tiene gran
importancia, pues permite distinguir dentro del enfoque flujos entre los conceptos de
solvencia y sostenibilidad. Esta última se conseguiría sólo cuando la estacionariedad se
alcanza sin necesidad de ajustes bruscos por parte de los agentes privados en sus
sendas temporales de consumo e inversión o bien, cuando no son necesarias medidas
de ajuste por parte de las autoridades económicas. Por tanto, la existencia de cambios
estructurales implicaría la no sostenibilidad de los desequilibrios exteriores, aún si
existe solvencia a largo plazo. En tercer lugar, consideramos la existencia de relaciones
de dependencia o correlación cruzada entre los países miembros del panel
considerado. Esta situación es esperable en áreas integradas económicamente como
es el caso de la OCDE. Su no consideración genera sesgos en los resultados que
pueden haber conducido a error en investigaciones anteriores. Por último, se analiza la
relación dinámica entre las dos variables de ajuste para los casos de Estados Unidos y
de España, de especial relevancia por la evolución y magnitud de sus desequilibrios
exteriores durante el período de análisis considerado.
Los resultados obtenidos señalan que individualmente sólo tres países no han
experimentado cambios estructurales durante el período analizado: Italia e Irlanda en
el caso del saldo de balanza por cuenta corriente y Alemania para el caso del saldo
acumulado de deuda exterior. En el resto de países se han identificado hasta un
máximo de cuatro discontinuidades que se corresponden con cambios institucionales
de importancia o con medidas de política económica (en su mayoría variaciones de
— 73 — tipo de cambio). Estos resultados indican que las intervenciones institucionales o,
alguna forma de ajuste brusco, se desencadena con el fin de poder mantener la
sostenibilidad de las cuentas exteriores de los países de la OCDE estudiados. Esta
evidencia empírica iría en contra de la capacidad autorreguladora del mercado para
alcanzar un “ajuste suave” y, por tanto, también en contra de la llamada “doctrina
Lawson” que aboga por un “laissez-faire”. Con todo, el proceso de integración
financiera creciente entre los países de la OCDE puede estar relajando la restricción
exterior de sus economías. De hecho, los resultados obtenidos indican la existencia de
dependencia estadística en cada sección cruzada del panel analizado. Una vez que se
consideran contrastes que engloban la posibilidad de cambios estructurales y
dependencia “cross-section” los resultados indican estacionariedad conjunta del panel
y, por tanto, la solvencia (si bien no la sostenibilidad) en el área en su conjunto.
Este resultado se confirma al analizar la dinámica del ajuste para el caso de España y
los Estados Unidos, países con severos desequilibrios exteriores pero que, una vez
considerados los cambios estructurales, alcanzan un ajuste estacionario para el
conjunto de periodos, excepto para España a partir de 1996. Este hecho estaría
indicando que la solvencia exterior se recupera tras shocks de un tamaño importante
sobre los saldos externos y que, sin embargo, el alto grado de globalización financiera
puede haber aumentado la persistencia de los desequilibrios en los miembros
periféricos de la Unión Monetaria Europea, como es el caso de España.
— 74 — NORMAS DE PUBLICACIÓN DE PAPELES DE TRABAJO DEL
INSTITUTO DE ESTUDIOS FISCALES
Esta colección de Papeles de Trabajo tiene como objetivo ofrecer un vehículo de
expresión a todas aquellas personas interasadas en los temas de Economía Pública. Las
normas para la presentación y selección de originales son las siguientes:
1. Todos los originales que se presenten estarán sometidos a evaluación y podrán
ser directamente aceptados para su publicación, aceptados sujetos a revisión, o
rechazados.
2. Los trabajos deberán enviarse por duplicado a la Subdirección de Estudios
Tributarios. Instituto de Estudios Fiscales. Avda. Cardenal Herrera Oria, 378. 28035
Madrid.
3. La extensión máxima de texto escrito, incluidos apéndices y referencias
bibliográfícas será de 7000 palabras.
4. Los originales deberán presentarse mecanografiados a doble espacio. En la primera
página deberá aparecer el título del trabajo, el nombre del autor(es) y la institución a la
que pertenece, así como su dirección postal y electrónica. Además, en la primera
página aparecerá también un abstract de no más de 125 palabras, los códigos JEL y las
palabras clave.
5. Los epígrafes irán numerados secuencialmente siguiendo la numeración arábiga.
Las notas al texto irán numeradas correlativamente y aparecerán al pie de la
correspondiente página. Las fórmulas matemáticas se numerarán secuencialmente
ajustadas al margen derecho de las mismas. La bibliografía aparecerá al final del
trabajo, bajo la inscripción “Referencias” por orden alfabético de autores y, en cada
una, ajustándose al siguiente orden: autor(es), año de publicación (distinguiendo a, b, c
si hay varias correspondientes al mismo autor(es) y año), título del artículo o libro,
título de la revista en cursiva, número de la revista y páginas.
6. En caso de que aparezcan tablas y gráficos, éstos podrán incorporarse
directamente al texto o, alternativamente, presentarse todos juntos y debidamente
numerados al final del trabajo, antes de la bibliografía.
7. En cualquier caso, se deberá adjuntar un disquete con el trabajo en formato word.
Siempre que el documento presente tablas y/o gráficos, éstos deberán aparecer en
ficheros independientes. Asimismo, en caso de que los gráficos procedan de tablas
creadas en excel, estas deberán incorporarse en el disquete debidamente identificadas.
Junto al original del Papel de Trabajo se entregará también un resumen
de un máximo de dos folios que contenga las principales implicaciones de
política económica que se deriven de la investigación realizada.
— 75 — PUBLISHING GUIDELINES OF WORKING PAPERS AT THE
INSTITUTE FOR FISCAL STUDIES
This serie of Papeles de Trabajo (working papers) aims to provide those having an
interest in Public Economics with a vehicle to publicize their ideas. The rules gover­
ning submission and selection of papers are the following:
1. The manuscripts submitted will all be assessed and may be directly accepted for
publication, accepted with subjections for revision or rejected.
2. The papers shall be sent in duplicate to Subdirección General de Estudios
Tributarios (The Deputy Direction of Tax Studies), Instituto de Estudios Fiscales
(Institute for Fiscal Studies), Avenida del Cardenal Herrera Oria, nº 378, Madrid
28035.
3. The maximum length of the text including appendices and bibliography will be no
more than 7000 words.
4. The originals should be double spaced. The first page of the manuscript should
contain the following information: (1) the title; (2) the name and the institutional affi­
liation of the author(s); (3) an abstract of no more than 125 words; (4) JEL codes and
keywords; (5) the postal and e-mail address of the corresponding author.
5. Sections will be numbered in sequence with arabic numerals. Footnotes will be
numbered correlatively and will appear at the foot of the corresponding page.
Mathematical formulae will be numbered on the right margin of the page in sequence.
Bibliographical references will appear at the end of the paper under the heading
“References” in alphabetical order of authors. Each reference will have to include in this
order the following terms of references: author(s), publishing date (with an a, b or c in
case there are several references to the same author(s) and year), title of the article or
book, name of the journal in italics, number of the issue and pages.
6. If tables and graphs are necessary, they may be included directly in the text or
alternatively presented altogether and duly numbered at the end of the paper, before
the bibliography.
7. In any case, a floppy disk will be enclosed in Word format. Whenever the
document provides tables and/or graphs, they must be contained in separate files.
Furthermore, if graphs are drawn from tables within the Excell package, these must
be included in the floppy disk and duly identified.
Together with the original copy of the working paper a brief two-page
summary highlighting the main policy implications derived from the
research is also requested.
— 77 — ÚLTIMOS PAPELES DE TRABAJO EDITADOS POR EL INSTITUTO DE ESTUDIOS FISCALES 2004
01/04 Una propuesta para la regulación de precios en el sector del agua: el caso español. Autores: M.a Ángeles García Valiñas y Manuel Antonio Muñiz Pérez. 02/04 Eficiencia en educación secundaria e inputs no controlables: sensibilidad de los
resultados ante modelos alternativos.
Autores: José Manuel Cordero Ferrera, Francisco Pedraja Chaparro y Javier Salinas Jiménez.
03/04 Los efectos de la política fiscal sobre el ahorro privado: evidencia para la OCDE.
Autores: Montserrat Ferre Carracedo, Agustín García García y Julián Ramajo Hernández.
04/04 ¿Qué ha sucedido con la estabilidad del empleo en España? Un análisis desagregado
con datos de la EPA: 1987-2003.
Autores: José María Arranz y Carlos García-Serrano.
05/04 La seguridad del empleo en España: evidencia con datos de la EPA (1987-2003).
Autores: José María Arranz y Carlos García-Serrano.
06/04 La ley de Wagner: un análisis sintético.
Autor: Manuel Jaén García.
07/04 La vivienda y la reforma fiscal de 1998: un ejercicio de simulación.
Autor: Miguel Ángel López García.
08/04 Modelo dual de IRPF y equidad: un nuevo enfoque teórico y su aplicación al caso español.
Autor: Fidel Picos Sánchez.
09/04 Public expenditure dynamics in Spain: a simplified model of its determinants.
Autores: Manuel Jaén García y Luis Palma Martos.
10/04 Simulación sobre los hogares españoles de la reforma del IRPF de 2003. Efectos sobre
la oferta laboral, recaudación, distribución y bienestar.
Autores: Juan Manuel Castañer Carrasco, Desiderio Romero Jordán y José Félix Sanz Sanz.
11/04 Financiación de las Haciendas regionales españolas y experiencia comparada.
Autor: David Cantarero Prieto.
12/04 Multidimensional indices of housing deprivation with application to Spain.
Autores: Luis Ayala y Carolina Navarro.
13/04 Multiple ocurrence of welfare recipiency: determinants and policy implications.
Autores: Luis Ayala y Magdalena Rodríguez.
14/04 Imposición efectiva sobre las rentas laborales en la reforma del impuesto sobre la renta
personal (IRPF) de 2003 en España.
Autoras: María Pazos Morán y Teresa Pérez Barrasa.
15/04 Factores determinantes de la distribución personal de la renta: un estudio empírico a
partir del PHOGUE.
Autores: Marta Pascual y José María Sarabia.
16/04 Política familiar, imposición efectiva e incentivos al trabajo en la reforma de la
imposición sobre la renta personal (IRPF) de 2003 en España.
Autoras: María Pazos Morán y Teresa Pérez Barrasa.
17/04 Efectos del déficit público: evidencia empírica mediante un modelo de panel dinámico
para los países de la Unión Europea.
Autor: César Pérez López.
— 79 — 18/04 Inequality, poverty and mobility: Choosing income or consumption as welfare
indicators.
Autores: Carlos Gradín, Olga Cantó y Coral del Río.
19/04 Tendencias internacionales en la financiación del gasto sanitario.
Autora: Rosa María Urbanos Garrido.
20/04 El ejercicio de la capacidad normativa de las CCAA en los tributos cedidos: una
primera evaluación a través de los tipos impositivos efectivos en el IRPF.
Autores: José María Durán y Alejandro Esteller.
21/04 Explaining. budgetary indiscipline: evidence from spanish municipalities.
Autores: Ignacio Lago-Peñas y Santiago Lago-Peñas.
22/04 Local governmets' asymmetric reactions to grants: looking for the reasons.
Autor: Santiago Lago-Peñas.
23/04 Un pacto de estabilidad para el control del endeudamiento autonómico.
Autor: Roberto Fernández Llera
24/04 Una medida de la calidad del producto de la atención primaria aplicable a los análisis
DEA de eficiencia.
Autora: Mariola Pinillos García.
25/04 Distribución de la renta, crecimiento y política fiscal.
Autor: Miguel Ángel Galindo Martín.
26/04 Políticas de inspección óptimas y cumplimiento fiscal.
Autores: Inés Macho Stadler y David Pérez Castrillo.
27/04 ¿Por qué ahorra la gente en planes de pensiones individuales?
Autores: Félix Domínguez Barrero y Julio López-Laborda.
28/04 L a reforma del Impuesto sobre Actividades Económicas: una valoración con
microdatos de la ciudad de Zaragoza.
Autores: Julio López-Laborda, M.ª Carmen Trueba Cortés y Anabel Zárate Marco.
29/04 Is an inequality-neutral flat tax reform really neutral?
Autores: Juan Prieto-Rodríguez, Juan Gabriel Rodríguez y Rafael Salas.
30/04 El equilibrio presupuestario: las restricciones sobre el déficit.
Autora: Belén Fernández Castro.
2005
01/05 Efectividad de la política de cooperación en innovación: evidencia empírica española.
Autores: Joost Heijs, Liliana Herrera, Mikel Buesa, Javier Sáiz Briones y Patricia Valadez.
02/05 A probabilistic nonparametric estimator.
Autores: Juan Gabriel Rodríguez y Rafael Salas.
03/05 Efectos redistributivos del sistema de pensiones de la seguridad social y factores
determinantes de la elección de la edad de jubilación. Un análisis por comunidades
autónomas.
Autores: Alfonso Utrilla de la Hoz y Yolanda Ubago Martínez.
14/05 La relación entre los niveles de precios y los niveles de renta y productividad en los países
de la zona euro: implicaciones de la convergencia real sobre los diferenciales de inflación.
Autora: Ana R. Martínez Cañete.
05/05 La Reforma de la Regulación en el contexto autonómico. Autor: Jaime Vallés Giménez. — 80 — 06/05 Desigualdad y bienestar en la distribución intraterritorial de la renta, 1973-2000.
Autores: Luis Ayala Cañón, Antonio Jurado Málaga y Francisco Pedraja Chaparro.
07/05 Precios inmobiliarios, renta y tipos de interés en España.
Autor: Miguel Ángel López García.
08/05 Un análisis con microdatos de la normativa de control del endeudamiento local.
Autores: Jaime Vallés Giménez, Pedro Pascual Arzoz y Fermín Cabasés Hita.
09/05 Macroeconomics effects of an indirect taxation reform under imperfect competition.
Autor: Ramón J. Torregrosa.
10/05 Análisis de incidencia del gasto público en educación superior: nuevas aproximaciones.
Autora: María Gil Izquierdo.
11/05 Feminización de la pobreza: un análisis dinámico.
Autora: María Martínez Izquierdo.
12/05 Efectos del impuesto sobre las ventas minoristas de determinados hidrocarburos en la
economía extremeña: un análisis mediante modelos de equilibrio general aplicado.
Autores: Francisco Javier de Miguel Vélez, Manuel Alejandro Cardenete Flores y Jesús
Pérez Mayo.
13/05 La tarifa lineal de Pareto en el contexto de la reforma del IRPF.
Autores: Luis José Imedio Olmedo, Encarnación Macarena Parrado Gallardo y María
Dolores Sarrión Gavilán.
14/05 Modelling tax decentralisation and regional growth.
Autores: Ramiro Gil-Serrate y Julio López-Laborda.
15/05 Interactions inequality-polarization: characterization results.
Autores: Juan Prieto-Rodríguez, Juan Gabriel Rodríguez y Rafael Salas.
16/05 Políticas de competencia impositiva y crecimiento: el caso irlandés.
Autores: Santiago Díaz de Sarralde, Carlos Garcimartín y Luis Rivas.
17/05 Optimal provision of public inputs in a second-best scenario.
Autores: Diego Martínez López y A. Jesús Sánchez Fuentes.
18/05 Nuevas estimaciones del pleno empleo de las regiones españolas.
Autores: Javier Capó Parrilla y Francisco Gómez García.
19/05 US deficit sustainability revisited: a multiple structural change approach.
Autores: Óscar Bajo-Rubio. Carmen Díaz-Roldán y Vicente Esteve.
20/05 Aproximación a los pesos de calidad de vida de los “Años de Vida Ajustados por
Calidad” mediante el estado de salud autopercibido.
Autores: Anna García-Altés, Jaime Pinilla y Salvador Peiró.
21/05 Redistribución y progresividad en el Impuesto sobre Sucesiones y Donaciones: una
aplicación al caso de Aragón.
Autor: Miguel Ángel Barberán Lahuerta.
22/05 Estimación de los rendimientos y la depreciación del capital humano para las regiones
del sur de España.
Autora: Inés P. Murillo.
23/05 El doble dividendo de la imposición ambiental. Una puesta al día.
Autor: Miguel Enrique Rodríguez Méndez.
24/05 Testing for long-run purchasing power parity in the post bretton woods era: evidence
from old and new tests.
Autor: Julián Ramajo Hernández y Montserrat Ferré Cariacedo.
— 81 — 25/05 Análisis de los factores determinantes de las desigualdades internacionales en las
emisiones de CO2 per cápita aplicando el enfoque distributivo: una metodología de
descomposición por factores de Kaya.
Autores: Juan Antonio Duro Moreno y Emilio Padilla Rosa.
26/05 Planificación fiscal con el impuesto dual sobre la renta.
Autores: Félix Domínguez Barrero y Julio López Laborda.
27/05 El coste recaudatorio de las reducciones por aportaciones a planes de pensiones y las
deducciones por inversión en vivienda en el IRPF 2002.
Autores: Carmen Marcos García, Alfredo Moreno Sáez, Teresa Pérez Barrasa y César
Pérez López.
28/05 La muestra de declarantes IEF-AEAT 2002 y la simulación de reformas fiscales:
descripción y aplicación práctica.
Autores: Alfredo Moreno, Fidel Picos, Santiago Díaz de Sarralde, María Antiqueira y
Lucía Torrejón.
2006
01/06 Capital gains taxation and progressivity.
Autor: Julio López Laborda.
02/06 Pigou’s dividend versus Ramsey’s dividend in the double dividend literature.
Autores: Eduardo L. Giménez y Miguel Rodríguez.
03/06 Assessing tax reforms. Critical comments and proposal: the level and distance effects.
Autores: Santiago Díaz de Sarralde Míguez y Jesús Ruiz-Huerta Carbonell.
04/06 Incidencia y tipos efectivos del impuesto sobre el patrimonio e impuesto sobre
sucesiones y donaciones.
Autora: Laura de Pablos Escobar.
05/06 Descentralización fiscal y crecimiento económico en las regiones españolas. Autores: Patricio Pérez González y David Cantarero Prieto. 16/06 Efectos de la corrupción sobre la productividad: un estudio empírico para los países
de la OCDE.
Autores: Javier Salinas Jiménez y M.ª del Mar Salinas Jiménez.
07/06 Simulación de las implicaciones del equilibrio presupuestario sobre la política de
inversión de las comunidades autónomas.
Autores: Jaime Vallés Giménez y Anabel Zárate Marco.
18/06 The composition of public spending and the nationalization of party sistems in western
Europe.
Autores: Ignacio Lago-Peñas y Santiago Lago.Peñas.
09/06 Factores explicativos de la actividad reguladora de las Comunidades Autónomas
(1989-2001).
Autores: Julio López Laborda y Jaime Vallés Giménez.
10/06 Disciplina credititicia de las Comunidades Autónomas.
Autor: Roberto Fernández Llera.
11/06 Are the tax mix and the fiscal pressure converging in the European Union?.
Autor: Francisco J. Delgado Rivero.
12/06 Redistribución, inequidad vertical y horizontal en el impuesto sobre la renta de las
personas físicas (1982-1998).
Autora: Irene Perrote.
— 82 — 13/06 Análisis económico del rendimiento en la prueba de conocimientos y destrezas
imprescindibles de la Comunidad de Madrid.
Autores: David Trillo del Pozo, Marta Pérez Garrido y José Marcos Crespo.
14/06 Análisis de los procesos privatizadores de empresas públicas en el ámbito internacional.
Motivaciones: moda política versus necesidad económica.
Autores: Almudena Guarnido Rueda, Manuel Jaén García e Ignacio Amate Fortes.
15/06 Privatización y liberalización del sector telefónico español.
Autores: Almudena Guarnido Rueda, Manuel Jaén García e Ignacio Amate Fortes.
16/06 Un análisis taxonómico de las políticas para PYME en Europa: objetivos, instrumentos
y empresas beneficiarias.
Autor: Antonio Fonfría Mesa.
17/06 Modelo de red de cooperación en los parques tecnológicos: un estudio comparado.
Autora: Beatriz González Vázquez.
18/06 Explorando la demanda de carburantes de los hogares españoles: un análisis de
sensibilidad.
Autores: Santiago Álvarez García, Marta Jorge García-Inés y Desiderio Romero Jordán.
19/06 Cross-country income mobility comparisons under panel attrition: the relevance of
weighting schemes.
Autores: Luis Ayala, Carolina Navarro y Mercedes Sastre.
20/06 Financiación Autonómica: algunos escenarios de reforma de los espacios fiscales.
Autores: Ana Herrero Alcalde, Santiago Díaz de Sarralde, Javier Loscos Fernández,
María Antiqueira y José Manuel Tránchez.
21/06 Child nutrition and multiple equilibria in the human capital transition function.
Autores: Berta Rivera, Luis Currais y Paolo Rungo.
22/06 Actitudes de los españoles hacia la hacienda pública.
Autor: José Luis Sáez Lozano.
23/06 Progresividad y redistribución a través del IRPF español: un análisis de bienestar social
para el periodo 1982-1998.
Autores: Jorge Onrubia Fernández, María del Carmen Rodado Ruiz, Santiago Díaz de
Sarralde y César Pérez López.
24/06 Análisis descriptivo del gasto sanitario español: evolución, desglose, comparativa
internacional y relación con la renta.
Autor: Manuel García Goñi.
25/06 El tratamiento de las fuentes de renta en el IRPF y su influencia en la desigualdad y la
redistribución.
Autores: Luis Ayala Cañón, Jorge Onrubia Fernández y María del Carmen Rodado Ruiz.
26/06 La reforma del IRPF de 2007: una evaluación de sus efectos.
Autores: Santiago Díaz de Sarralde Míguez, Fidel Picos Sánchez, Alfredo Moreno Sáez,
Lucía Torrejón Sanz y María Antiqueira Pérez.
27/06 Proyección del cuadro macroeconómico y de las cuentas de los sectores
institucionales mediante un modelo de equilibrio.
Autores: Ana María Abad, Ángel Cuevas y Enrique M. Quilis.
28/06 Análisis de la propuesta del tesoro Británico “Fiscal Stabilisation and EMU” y de sus
implicaciones para la política económica en la Unión Europea.
Autor: Juan E. Castañeda Fernández.
— 83 — 29/06 Choosing to be different (or not): personal income taxes at the subnational level in
Canada and Spain.
Autores: Violeta Ruiz Almendral y François Vaillancourt.
30/06 A projection model of the contributory pension expenditure of the Spanish social
security system: 2004-2050.
Autores: Joan Gil, Miguel Ángel Lopez-García, Jorge Onrubia, Concepció Patxot y
Guadalupe Souto.
2007
11/07 Efectos macroeconómicos de las políticas fiscales en la UE. Autores: Oriol Roca Sagalés y Alfredo M. Pereira. 02/07 Deficit sustainability and inflation in EMU: an analysis from the fiscal theory of the
price level.
Autores: Óscar Bajo-Rubio, Carmen Díaz-Roldán y Vicente Esteve.
03/07 Contraste empírico del modelo monetario de tipos de cambio: cointegración y ajuste
no lineal.
Autor: Julián Ramajo Hernández.
04/07 An empirical analysis of capital taxation: equity vs. tax compiance. Autores: José M.a Durán Cabré y Alejandro Esteller Moré. 05/07 Education and health in the OECD: a macroeconomic approach. Autoras: Cecilia Albert y María A. Davia. 06/07 Understanding the effect of education on health across European countries. Autoras: Cecilia Albert y María A. Davia. 07/07 Polarization, fractionalization and conflict. Autores: Joan Esteban y Debraj Ray. 08/07 Immigration in a segmented labor market: the effects on welfare. Autor: Javier Vázquez Grenno. 09/07 On the role of public debt in an OLG Model with endogenous labor supply. Autor: Miguel Ángel López García. 10/07 Assessing profitability in rice cultivation using the Policy Matrix Analysis and profit­
efficient data.
Autores: Andrés J. Picazo-Tadeo, Ernest Reig y Vicent Estruch.
11/07 Equidad y redistribución en el Impuesto sobre Sucesiones y Donaciones: análisis de los
efectos de las reformas autonómicas.
Autores: Miguel Ángel Barberán Lahuerta y Marta Melguizo Garde.
12/07 Valoración y determinantes del stock de capital salud en la Comunidad Canaria y Cataluña.
Autores: Juan Oliva y Néboa Zozaya.
13/07 La nivelación en el marco de la financiación de las Comunidades Autónomas.
Autores: Ana Herrero Alcalde y Jorge Martínez-Vázquez.
14/07 El gasto en defensa en los países desarrollados: evolución y factores explicativos.
Autor: Antonio Fonfría Mesa.
15/07 Los costes del servicio de abastecimiento de agua. Un análisis necesario para la
regulación de precios.
Autores: Ramón Barberán Ortí, Alicia Costa Toda y Alfonso Alegre Val.
16/07 Precios, impuestos y compras transfronterizas de carburantes.
Autores: Andrés Leal Marcos, Julio López Laborda y Fernando Rodrigo Sauco.
— 84 —
17/07 Análisis de la distribución de las emisiones de CO2 a nivel internacional mediante la
adaptación del concepto y las medidas de polarización.
Autores: Juan Antonio Duro Moreno y Emilio Padilla Rosa.
18/07 Foreign direct investment and regional growth: an analysis of the Spanish case.
Autores: Óscar Bajo Rubio, Carmen Díaz Mora y Carmen Díaz Roldán.
19/07 Convergence of fiscal pressure in the EU: a time series approach.
Autores: Francisco J. Delgado y María José Presno.
20/07 Impuestos y protección medioambiental: preferencias y factores.
Autores: María de los Ángeles García Valiñas y Benno Torgler.
21/07 Modelización paramétrica de la distribución personal de la renta en España. Una
aproximación a partir de la distribución Beta generalizada de segunda especie.
Autores: Mercedes Prieto Alaiz y Carmelo García Pérez.
22/07 Desigualdad y delincuencia: una aplicación para España.
Autores: Rafael Muñoz de Bustillo, Fernando Martín Mayoral y Pablo de Pedraza.
23/07 Crecimiento económico, productividad y actividad normativa: el caso de las Comunidades
Autónomas.
Autor: Jaime Vallés Giménez.
24/07 Descentralización fiscal y tributación ambiental. El caso del agua en España.
Autores: Anabel Zárate Marco, Jaime Vallés Giménez y Carmen Trueba Cortés.
25/07 Tributación ambiental en un contexto federal. Una aplicación empírica para los
residuos industriales en España.
Autores: Anabel Zárate Marco, Jaime Vallés Giménez y Carmen Trueba Cortés.
26/07 Permisos de maternidad, paternidad y parentales en Europa: algunos elementos para
el análisis de la situación actual.
Autoras: Carmen Castro García y María Pazos Morán.
27/07 ¿Quién soporta las cotizaciones sociales empresariales?. Una panorámica de la literatura
empírica.
Autor: Ángel Melguizo Esteso.
28/07 Una propuesta de financiación municipal.
Autores: Manuel Esteban Cabrera y José Sánchez Maldonado.
29/07 Do R&D programs of different government levels overlap in the European Union.
Autoras: Isabel Busom y Andrea Fernández-Ribas.
30/07 Proyecciones de tablas de mortalidad dinámicas de España y sus Comunidades
Autónomas.
Autores: Javier Alonso Meseguer y Simón Sosvilla Rivero.
2008
11/08 Estudio descriptivo del voto económico en España. Autores: José Luis Sáez Lozano y Antonio M. Jaime Castillo. 12/08 The determinants of tax morale in comparative perspective: evidence from a multilevel
analysis.
Autores: Ignacio Lago-Peñas y Santiago Lago-Peñas.
13/08 Fiscal decentralization and the quality of government: evidence from panel data. Autores: Andreas P. Kyriacou y Oriol Roca-Sagalés. 14/08 The effects of multinationals on host economies: A CGE approach. Autores: María C. Latorre, Oscar Bajo-Rubio y Antonio G. Gómez-Plana.
— 85 — 15/08 Measuring the effect of spell recurrence on poverty dynamics. Autores: José María Arranz y Olga Cantó. 16/08 Aspectos distributivos de las diferencias salariales por razón de género en España: un
análisis por subgrupos poblacionales.
Autores: Carlos Gradín y Coral del Río.
17/08 Evaluating the regulator: winners and losers in the regulation of Spanish electricity
distribution (1988-2002).
Autores: Leticia Blázquez Gómez y Emili Grifell-Tatjé.
18/08 Interacción de la política monetaria y la política fiscal en la UEM: tipos de interés a
corto plazo y déficit público.
Autores: Jesús Manuel García Iglesias y Agustín García García.
19/08 A selection model of R&D intensity and market structure in Spanish forms.
Autor: Joaquín Artés.
10/08 Outsourcing behaviour: the role of sunk costs and firm and industry characteristics.
Autoras: Carmen Díaz Mora y Angela Triguero Cano.
11/08 How can the decommodified security ratio assess social protection systems?.
Autor: Georges Menahem.
12/08 Pension policies and income security in retirement: a critical assessment of recent
reforms in Portugal.
Autora: Maria Clara Murteira.
13/08 Do unemployment benefit legislative changes affect job finding? Evidence from the
Spanish 1992 UI reform act.
Autores: José M. Arranz, Fernando Muñoz Bullón y Juan Muro.
14/08 Migraciones interregionales en España y su relación con algunas políticas públicas.
Autora: María Martínez Torres.
15/08 Entradas y salidas de la pobreza en la Unión Europea: factores determinantes.
Autores: Guillermina Martín Reyes, Elena Bárcena Martín, Antonio Fernández Morales
y Antonio García Lizana.
16/08 Income mobility and economic inequality from a regional perspectiva.
Autores: Juan Prieto Rodríguez, Juan Gabriel Rodríguez y Rafael Salas.
17/08 A note on the use of calendar regressors.
Autor: Leandro Navarro Pablo.
18/08 Asimetrías y efectos desbordamiento en la transmisión de la política fiscal en la Unión
Europea: evidencia a partir de un enfoque VAR estructural.
Autor: Julián Ramajo.
19/08 Institutionalizing uncertainty: the choice of electoral formulas.
Autores: Gonzalo Fernández de Córdoba y Alberto Penadés.
20/08 A field experiment to study sex and age discrimination in selection processes for staff
recruitment in the Spanish labor market.
Autores: Rocío Albert, Lorenzo Escot, y José A. Fernández-Cornejo.
21/08 Descentralización y tamaño del sector público regional en España.
Autor: Patricio Pérez.
22/08 Multinationals and foreign direct investment: main theoretical strands and empirical
effects.
Autora: María C. Latorre.
— 86 — 23/08 Una aproximación no lineal al análisis del impacto de las finanzas públicas en el
crecimiento económico de los países de la UE-15, 1965-2007.
Autor: Diego Romero Ávila.
24/08 Consolidación y reparto de la base imponible del Impuesto sobre Sociedades entre los
Estados Miembros de la Unión Europea: consecuencias para España.
Autores: Félix Domínguez Barrero y Julio López Laborda.
25/08 La suficiencia dinámica del modelo de financiación autonómica en España, 2002-2006.
Autores: Catalina Barceló Maimó, María Marquès Caldentey y Joan Rosselló Villalonga.
26/08 Ayudas públicas en especie y en efectivo: justificaciones y aspectos metodológicos.
Autores: Laura Piedra Muñoz y Manuel Jaén García.
27/08 Las ayudas públicas al alquiler de la vivienda. un análisis empírico para evaluar sus
beneficios y costes.
Autores: Laura Piedra Muñoz y Manuel Jaén García.
28/08 Decentralization and spatial distribution of regional ecomonic activity: does equalization
matter?.
Autores: Santiago Lago-Peñas y Diego Martínez-López.
29/08 Childcare costs and Spanish mothers’s labour force participation.
Autora: Cristina Borra.
30/08 Pro-poor economic growth, inequality and fiscal policy: the case of Spanish regions.
Autores: Luis Ayala y Antonio Jurado.
2009
01/09 Does the balance of payments constrain economic growth?. Some evidence for the
new EU members.
Autores: Oscar Bajo-Rubio y Carmen Díaz-Roldán.
02/09 Imputación a valor de mercado de los rendimientos de la vivienda en Propiedad del
IRPF.
Autores: Luis Ayala Cañón, Jorge Onrubia Fernández y María del Carmen Rodado Ruiz.
03/09 Income poverty and multidimensional deprivation: lessons from cross-regional analysis.
Autores: Luis Ayala Cañón, Antonio Jurado y Jesús Perez-Mayo.
04/09 Reglas fiscales activas: el caso de España (1981-2007).
Autor: Juan E. Castañeda Fernández.
05/09 Índices trimestrales de volumen encadenados, ajuste estacional y Bechmarking.
Autores: Ana M.ª Abad, Ángel Cuevas y Enrique M. Quilis.
06/09 Fiscal decentralization and economic growth in OECD countries: matching spending
wit revenue decentralization.
Autores: Norman Gemmell, Richard Kneller e Ismael Sanz.
07/09 Una estimación del voto estratégico en las elecciones generales españolas, 2000-2008.
Autores: Enrique García Viñuela y Joaquín Artés.
08/09 La tributación del transporte como instrumento frente al cambio climático.
Autor: Miguel Buñuel González
09/09 The ins and outs of unemployment and the assimilation of recent immigrants in Spain.
Autores: José I. Silva y Javier Vázquez.
10/09 Decomposing the determinants of health care expenditure: the case of Spain.
Autores: David Cantarero Prieto y Santiago Lago-Peña.
— 87 — 11/09 La clase beta de medidas de desigualdad.
Autores: Luis José Imedio Olmedo, Elena Bárcena Martín y Encarnación M. Parrado
Gallardo.
12/09 Right incentives to enhance efficiency in public expenditure.
Autor: Tamón A. Takahashi Iturriaga.
13/09 Fiscal decentralization and public sector employment: a cross-country analysis.
Autores: Jorge Martínez-Vázquez y Ming-Hung Yao.
14/09 Factores explicativos de los resultados de las Comunidades Autónomas Españolas en
PISA 2006.
Autores: José Manuel Cordero Ferrera, Eva Crespo Cebada y Daniel Santín González.
15/09 A proposal to empirically evaluate the sensitivity of the speed of convergence in the EU.
Autoras: Sonia de Lucas Santos, Inmaculada Álvarez Ayuso y M.ª Jesús Delgado Rodríguez.
16/09 An assessment of the sustainability of current account imbalances in OECD countries.
Autores: Mariam Camarero, Josep Lluís Carrion-i-Silvestre y Cecilio Tamarit.
— 88 —