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Abstract

Recent theoretical and empirical analyses of the relation between the current account and the government budget balance suggest that the “twin deficits” relation is subject to structural changes. Most previous empirical analyses impose the change point without resorting to econometric testing. In this paper we utilize time series data to evaluate the impact of structural breaks on the long- and short-run relation between current account, government balance and investment in 22 OECD countries. We found that when allowing for the possible existence of structural breaks of unknown date, the data reveal more clearly the long-run relation between the current account and its determinants. Moreover, the empirical results show that the degree of financial integration is generally increasing in most OECD countries, including the leading non-EU economies. This contrasts some recent evidence on the persistence of the so-called Feldstein–Horioka puzzle.

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Notes

  1. According to the “current account targeting” hypothesis (Summers 1988) the government of a country may resort to fiscal policy to adjust its external position: this leads to a reverse causality nexus, running from the current account to the budget deficit.

  2. This led some authors to investigate the so called “twin divergence” phenomenon (Kim and Roubini 2003).

  3. Remark that \( ca_{t} = s_{t} - i_{t} \) (where the index t now refers to time), and since in the time series domain Eq. (3) becomes \( i_{t} = \alpha + \beta s_{t} + u_{t} \), it follows that \( ca_{t} = - \alpha + {\left( {1 - \beta } \right)}s_{t} - u_{t} \).

  4. See for instance Gokhale et al. (2001) for savings dynamic in postwar United States.

  5. Let \( \widehat{u}_{t} \) be the OLS residuals of Eq. (6). The auxiliary regression of the CRADF test is \( \Delta \widehat{u}_{t} = \gamma \widehat{u}_{{t - 1}} + {\sum\nolimits_{j = 1}^p {\Delta \widehat{u}_{{t - j}} + \eta _{t} } } \), where η t is a white noise. Starting from p = 5 we reduced the model until the Q statistics of the auxiliary regression residuals did not signal the presence of autocorrelation.

  6. The order of lags in the auxiliary ADF regression was selected as explained in the preceding footnote 4.

  7. We utilized as a trimming parameter 5, corresponding to about 11% of the available observations.

  8. In the case of Canada the I(0) variable i t is omitted from the long-run regression.

  9. The private savings ratio increased sharply by about seven GDP points from the end of the Sixties to the mid-Seventies.

  10. The endogenously determined change point dates for the long-run relation fall respectively in 1980 and 1994, rather than in 1989.

  11. The ADF* statistics for Canada however are unable to reject the null of non-cointegration.

  12. The cointegration test in the 1975–2005 subsample was performed using the small sample critical values of Blangiewicz and Charemza (1990).

  13. See for instance Blanchard and Giavazzi (2002) and Fidrmuc (2003).

  14. For the US we estimate coefficients of investment around –1, which contrasts the evidence in Obstfeld and Rogoff (1995), as well as the estimated −0.09 in Fidrmuc (2003); the coefficient of investment in Japan goes from zero to −0.45 after the structural break in 1981.

  15. We excluded from the sample due to lack of data the following countries: Czech Republic, Hungary, Iceland, Korea, Luxembourg, Mexico, Poland and Slovak Republic.

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Correspondence to Alberto Bagnai.

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I thank two anonymous referees for the many useful suggestions. Eleonora Pierucci supplied valuable research assistance.

Appendix: data sources

Appendix: data sources

Our data sample consists of annual data running from 1960 to 2005 (data for 2005 are the latest OECD projections) for 22 OECD countries.Footnote 15 The main data source is the 2004#2 CD-ROM edition of the “OECD Economic Outlook Statistics and Projections”. The Economic Outlook codes for the three variables in our dataset are as follows ca t : CBGDPR; \( s^{{\text{G}}}_{t} \): NLGQ; i t : 100 × ITV / GDPV.

In some cases missing values of the current account balance ca t were approximated by the trade balance, reconstructed as 100 × (XGSV − MGSV) / GDPV. This applies to Austria (1960–1969), Belgium (1960–1974), Denmark (1960–1974), Finland (1960–1974), France (1960–1974), Greece (1960–1974), Ireland (1960–1974), Italy (1960–1970), Japan (1960–1984), Netherlands (1960–1974), Norway (1960–1974), New Zealand (1961–1970), Portugal (1960–1974), Spain (1960–1974), Turkey (1967–1974), Sweden (1960–1974).

Supplementary data sources were utilized when the time series in OECD (2004) did not start in 1960.

The IFS 2004#9 CD-ROM edition was utilized for reconstructing the time series of the trade balance in: Switzerland (1960–1971), and the time series of \( s^{{\text{G}}}_{t} \) (defined as the ratio of the general government balance, IFS series code 80..ZF or 80G.ZF, to the GDP, IFS series code 99B.ZF...) in the following cases: Austria (1960–1963), Canada (1960–1980), Ireland (1960–1976), Netherlands (1960–1968), Norway (1960–1974), New Zealand (1960–1986), Switzerland (1960–1989), Turkey (1967–1978).

The 1999#1 edition of the OSC was utilized for reconstructing the time series of i t for Denmark (1960–1965).

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Bagnai, A. Structural breaks and the twin deficits hypothesis. IEEP 3, 137–155 (2006). https://doi.org/10.1007/s10368-006-0050-8

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