Abstract
This paper investigates the issue of monetary interdependence among members of the European Monetary System over the period 1979–91 and the leadership role attributed to the German central bank in the process of monetary integration, and looks for possible changes in central banks’ behaviors. Econometric evidence supports somewhat the German leadership role but suggests also the development of an increased interdependence between French and German monetary policies after 1981–82; meanwhile, the Italian and more recently the Spanish central banks appear to have preserved a more significant measure of monetary autonomy.
I. Introduction
The process of European monetary integration that followed the collapse of the Bretton Woods system in the early seventies with the creation of “the Snake” took its full fledged form in March 1979 with the development of the European Monetary System. This system is endowed with an extensive mechanism of financial support and an Exchange Rate Mechanism to contain the fluctuations of the participating currencies within margins of 2.25 percent—6 percent for new members at the initial stage—around bilateral central rates whose possible adjustments are subject to mutual agreement. The EMS was further enhanced by the 1987 Nyborg Agreement which allowed a more liberal use of the financial support and by the 1991 Maastricht Agreement which called for European Monetary Union by the end of the century.
Efforts made by participating central banks in the EMS context since 1979 to stabilize their exchange rates have led to a gradual convergence of their nominal interest rates and a progressive loss of monetary independence. It has been argued, however, that the burden imposed by the convergence process on the conduct of monetary policy was not equally shared among EMS members and that the German central bank assumed a leadership role during this period. This paper reviews this issue by investigating possible changes in the conduct of monetary policy in EC central banks following their participation in the ERM, and the rationale underlying these changes.
The degree of monetary dependence between ERM central banks is appreciated by comparing the fluctuations of their intervention rates or most closely related short-term domestic interest rates through three avenues: (i) Granger causality tests, (ii) causality tests including an error correction term determined in a co-integration framework, and (iii) reaction functions embedded in an error correction model. Possible changes in regime for these tests were determined endogenously in contrast with previous studies. The results of these tests suggest that the French, German and Italian central banks did not alter the implementation of monetary policy from the inception of the EMS but only after 1981–82. After this period, there is evidence of an increase interdependence between their interest rates, in particular between France and Germany. However, the fact that the German interest rate appears also to respond strongly to the variations of the U.S. dollar, which is not the case for other ERM members, provides some support to the leadership thesis. In sum, the dominance of the German central bank was not exclusive of an increased interdependence between ERM central banks, including that of Germany. The main reasons accounting for changes in regime are related to technical changes in the implementation of monetary policy and to the introduction of adjustment programs, generally in the wake of a major realignment, carried out to dampen inflation with a view to strengthening the credibility of the stabilization of the exchange rate in the ERM band. The paper is structured as follows: section II discusses the pro and con arguments for the German leadership thesis; section III describes the methodology of the econometric tests; section IV reports empirical results; Section V concludes.
II. The European Monetary Convergence Process: The Two Views
It has been argued that the process of European monetary integration, which developed under the EMS, took place under the leadership of the German monetary authorities in an asymmetrical context. The Bundesbank, viewed by the market as the conservative central bank, determines its monetary policy and intervention rates according either to domestic objectives or to changes in interest rates or exchange rates of the main non-ERM currencies, namely the U.S. Dollar. Meanwhile, other ERM central banks adjust their interest rates so as to stabilize the exchange rate of their currency against the deutsche mark, thus preserving the credibility of their anti-inflationary stance. There is an abundant theoretical and empirical literature in favor of this analysis (Melitz (1986), Giavazzi and Giovannini (1987), Fisher (1987), Artus and alii (1991). Also, the 1991–92 increases in the Bundesbank intervention rates, followed by similar moves in most of the other ERM central banks in particular at end 1991, provided a timely apparent support to the German leadership thesis.
This view has been challenged on theoretical and institutional grounds. On the theoretical side, Von Hagen and Fratianni (1990 a) contested the conceptual foundations of the credibility argument. To the extent that, in an exchange rate arrangement, the cost of inflating is lower for the largest countries than for the smallest (since the largest members can more easily export their inflation), the Bundesbank has an incentive to inflate, inducing higher inflation expectations and a higher German equilibrium inflation rate. On the institutional side, several features provided for a symmetrical functioning of the ERM, while others allowed central banks to maintain a measure of monetary independence. Factors of symmetry are embedded in the characteristics of the ERM. First, fluctuation margins are defined in terms of bilateral central rates for all participating currencies identically. This is in contrast to the Bretton Woods system in which parities and margins were defined only with reference to the U.S. Dollar, which meant that the maximum fluctuation between two currencies—excluding the U.S. Dollar—was twice the margins vis-a-vis the U.S. Dollar. Second, foreign exchange interventions at the margins affect the monetary base of the two intervening countries as the interventions by the central bank under pressure are financed by resources provided by other ERM central banks through the Very Short-Term Monetary Mechanism. 5/ Therefore, the contractionary monetary impact in the former economy is matched by an expansionary impact in the rest of the ERM countries that tend to restore the equality of interest rates among ERM members. While these elements were providing conditions of symmetry in the functioning of the ERM, others allowed ERM central banks to preserve a significant measure of monetary autonomy. Some features were temporary in nature while others were imbedded in the ERM. Capital controls, at least up to July 1990 for the core ERM economies, were used in some countries to sever the link between the internal and external money markets of the domestic currency. Market segmentation translated at times into large discrepancies between domestic and Euro-interest rates as in the case of the French franc or the Italian lira. The effect of imperfect capital mobility might also have been compounded by imperfect capital substitutability allowing central banks to stabilize the exchange rate, at least temporarily, through sterilized intervention, thus preserving their domestic monetary stance. Some exchange rate flexibility was also preserved, first within the fluctuation margins particularly important for currencies operating within the broad band, and second through successive exchange rate realignments, at least up to 1987. The following table points out that the long term drift of the French franc against the deutsche mark in the first nine years of the EMS was hardly different from the depreciation registered over 1973–78.
Average Annual Depreciation against Deutsche Mark
(In percent)
Average Annual Depreciation against Deutsche Mark
(In percent)
| Period | French franc | lira |
|---|---|---|
| 1973–78 | 5.6 | 15.1 |
| 1978–87 | 4.4 | 5.6 |
| 1987–91 | 0.4 | 0.7 |
Average Annual Depreciation against Deutsche Mark
(In percent)
| Period | French franc | lira |
|---|---|---|
| 1973–78 | 5.6 | 15.1 |
| 1978–87 | 4.4 | 5.6 |
| 1987–91 | 0.4 | 0.7 |
The role of these factors was illustrated in various empirical studies. De Grauwe (1989) insisted on the role of capital market segmentation, and Von Hagen and Fratianni (1990 b) provided evidence that “Germany’s policy is not independent from other members’ policies in the EMS” and that the relative independence of the Bundesbank has been significantly curtailed after March 1983. In the same vein, Gardner and Perraudin (1991) found that with the apparent weakening of the anti-inflationary thrust of German financial policies in the wake of German unification in 1990, the German leadership role in the implementation of monetary policy in other ERM countries was reduced while interest rates in Germany appeared more strongly affected by French interest rates.
In the following sections, the issues of the German leadership and degree of monetary autonomy enjoyed by other ERM members is reviewed with an attempt to determine endogenously any possible change in regime, in contrast with previous papers in which the change in regime is exogenously imposed.
III. Methodology
The measure of the degree of monetary interdependence between ERM countries raises two methodological issues: (i) the determination of the relevant instrument and objective targeted by the authorities, and (ii) the specification of the implicit reaction function used by the authorities to achieve their target.
In a strictly fixed exchange rate system, the stabilization of the exchange rate provides the final objective of monetary policy; monetary independence is lost and interest rates in participating countries adjust to a common level. In the ERM, as indicated above, the leeway offered by the system allowed central banks to stabilize their exchange rate in the short term through sterilized intervention and to manage other monetary instruments to account both for domestic developments and increasing international interest rate linkages as the world capital markets became more integrated. This phenomenon, in particular, has led central banks in all ERM countries, including Germany and more recently Spain, to put more emphasis on the close monitoring of interest rates rather than on the control of bank liquidity. Therefore, to test for the degree of central bank independence in the ERM, it seems appropriate to look for linkages between their interest rates. Among the various available interest rates—domestic and Euro-interest rates with short- or long-term maturities—the most relevant appears to be the short-term domestic interest rates with maturity ranging from day-to-day to three-month. Two main reasons account for this choice. First, in the context of segmented money markets experienced by most ERM countries during the period under review, domestic interest rates provide a more accurate representation of ERM central banks’ behavior than Euro-interest rates which were at times strongly tainted by expectations of impending exchange rate realignments in the EMS. Second, short term maturities appear more appropriate, as ERM central banks intervened only on the lower end of the yield curve. 3/
Before estimating the assumed reaction function on the domestic interest rate, two kinds of Granger causality tests were carried out with respect to other ERM interest rates to determine the respective importance of short- and long-term dynamics in ERM central banks’ reaction functions. The first causality test (Granger 1969, Sargent 1976), applied to nominal interest rates, is interested only in the short term “causal” linkages between two series. It determines the extent to which the knowledge of the present and lagged values of a given explanatory series improves on the prediction of an endogenous series based only on its own lagged values. A maximum lag of three months was selected. The second causality test (Granger (1988)) investigates the “causal” relationships in a co-integrated framework, combining short-term forecastability as in the previous test with the possible existence of a co-integrating relationship between the two series involved. Existence of a co-integrating relationship means that there exists a long-term equilibrium between the two series and that, for this to be possible, “there must be some causation between them to provide the necessary dynamics.” This approach was carried out by comparing the domestic money market rate of a given country to the German money market rate corrected by their inflation differentials and vice versa. The recognition of co-integration could provide evidence that ERM central banks adjusted their interest rate according to German interest rate and inflation performance, giving further support to the German leadership thesis. 4/
The outcome of the Granger causality tests was used to select the main explanatory variables in the reaction functions. The specification of the reaction function was completed by adding up to the previous causal relationships short-term indicators that appeared to have been representative of monetary tensions in the ERM bands, mainly foreign exchange interventions, deviations from central rates and the U.S. dollar fluctuations against the deutsche mark. Attempts to enter domestic indicators other than inflation rates failed to provide significant results. In the estimation technique, no attempt was made to account for possible simultaneity in the determination of foreign exchange interventions and exchange rates in particular. This should not bias significantly the estimates to the extent that most interventions were sterilized and could not accordingly stabilize lastingly exchange rates, in spite of market imperfections.
In order to test for any possible change in regime due to the weakening of the German leadership related either to the removal of capital controls and increased integration of European financial markets or to German unification or other factors, the Log-likelihood function proposed by Quandt (1958) was estimated for all above-mentioned tests and reaction functions. The relevant date for which there is a deterministic change in regime is the value which maximizes this Log-likelihood function. 2/
IV. Empirical Results
1. Short-term causality
Table 1 reports the results of the Granger causality test applied to determine the respective influence of the German interest rate on those of France, Italy, Spain, and the United States, and conversely of those rates on the German interest rate. The test is implemented in two steps. First, after taking the first differences of the series, 6/ each interest rate—excluding the German rate—is regressed on its own lagged values and on the German interest rate lagged values to test that the German rate does not cause the given interest rate, i.e., all coefficients on lagged German interest rates should not be significantly different from zero. Second, the German interest rate is regressed on its own lagged values and lagged values of the given interest rate to test that this rate does not cause the German interest rate, i.e., all coefficients on lagged values of the given interest rate should not be significantly different from zero. The results are presented for the two forms of the Granger causality tests: (i) simple causality in which only enter the lagged values of the exogenous variable, providing evidence of an unidirectional causality from the exogenous to the endogenous variable; and (ii) instantaneous causality in which the contemporaneous value of the exogenous variable is included, providing evidence of simultaneous influence. The most significant causalities emerging from these tests are those linking the variations of the French and italian interest rates to the past (simple causality) or present (instantaneous causality) variations of the German interest rate. Conversely, the variations of the German interest rate appear to have been caused to a significant extent by those of the U.S. interest rate while the variation of the latter are better accounted for by its own lagged values. These results seem to provide support for the German leadership thesis in the ERM. However, evidence of a strong instantaneous causality between the French and German interest rates suggests the existence of reciprocal influence between German and French monetary policies. This interpretation should be taken with caution, however, since the monthly periodicity of the series is not appropriate to disentangle the nature of a decision process which involved daily concertations between the two central banks.
Granger Casuality Tests on Short-term Domestic Interest Rates Expressed in First Differences 1/: Test of Two Regimes Over the period July 1979 to june 1991 2/
Regression of
The interest rates of the various currencies are represented by the following symbols: FRF, French franc; DEM, Deutsche mark; ITL, lira; PTA, peseta; USD, U.S. dollar.
The breaking points corresponding to changes in regime, as well as their degree of significance are presented in Table 2.
One and two asterisks indicate that X “Granger-causae” Y at levels of significance of 5 and 1 percent, respectively. Numbers in parentheses underlying F-statistics correspond to the degrees of freedom of the relevant test.
Granger Casuality Tests on Short-term Domestic Interest Rates Expressed in First Differences 1/: Test of Two Regimes Over the period July 1979 to june 1991 2/
Regression of
| F - Statistic on All βS=0 3/ | |||||||
|---|---|---|---|---|---|---|---|
| ΔY | ΔX | Simple Causality (j=1) | Instantaneous Causality (j=0) | ||||
| OVERALL PERIOD | REGIME 1 | REGIME 2 | OVERALL PERIOD | REGIME 1 | REGIME 2 | ||
| FRF | DEM | 5.46** (2.139) | 2.01 (2,33) | 5.40** (2,101) | 10.29** (2,138) | 3.96* (2,32) | 8.63** (2,100) |
| DEM | FRF | 1.54 (2,139) | 1.58 (2,18) | 0.86 (2,116) | 6.15** (2,138) | 3.67* (2,17) | 2.25 (2,115) |
| ITL | DEM | 4.86** (2,139) | 7.89** (2,76) | 2.12 (2,58) | 4.89** (2,138) | 8.23** (2,75) | 3.73* (2,57) |
| DEM | ITL | 0.19 (2,139) | 0.14 (2,18) | 0.17 (2,116) | 0.25 (2,138) | 0.36 (2,17) | 0.81 (2,15) |
| PTA | DEM | 3.39* (2,139) | 2.23 (2,54) | 0.41 (2,80) | 3.92* (2,138) | 2.47 (2,53) | 0.49 (2,79) |
| DEM | PTA | 0.42 (2,139) | 0.22 (2,18) | 0.22 (2,116) | 0.95 (2,138) | 0.47 (2,17) | 0.29 (2,115) |
| USD | DEM | 0.46 (2,139) | 0.12 (2,135) | 0.44 (2,99) | 1.30 (2,138) | 0.36 (2,34) | 1.35 (2,98) |
| DEM | USD | 4.75** (2,139) | 0.93 (2,18) | 3.12* (2,116) | 5.61** (2,138) | 0.89 (2,17) | 4.12* (2,115) |
The interest rates of the various currencies are represented by the following symbols: FRF, French franc; DEM, Deutsche mark; ITL, lira; PTA, peseta; USD, U.S. dollar.
The breaking points corresponding to changes in regime, as well as their degree of significance are presented in Table 2.
One and two asterisks indicate that X “Granger-causae” Y at levels of significance of 5 and 1 percent, respectively. Numbers in parentheses underlying F-statistics correspond to the degrees of freedom of the relevant test.
Granger Casuality Tests on Short-term Domestic Interest Rates Expressed in First Differences 1/: Test of Two Regimes Over the period July 1979 to june 1991 2/
Regression of
| F - Statistic on All βS=0 3/ | |||||||
|---|---|---|---|---|---|---|---|
| ΔY | ΔX | Simple Causality (j=1) | Instantaneous Causality (j=0) | ||||
| OVERALL PERIOD | REGIME 1 | REGIME 2 | OVERALL PERIOD | REGIME 1 | REGIME 2 | ||
| FRF | DEM | 5.46** (2.139) | 2.01 (2,33) | 5.40** (2,101) | 10.29** (2,138) | 3.96* (2,32) | 8.63** (2,100) |
| DEM | FRF | 1.54 (2,139) | 1.58 (2,18) | 0.86 (2,116) | 6.15** (2,138) | 3.67* (2,17) | 2.25 (2,115) |
| ITL | DEM | 4.86** (2,139) | 7.89** (2,76) | 2.12 (2,58) | 4.89** (2,138) | 8.23** (2,75) | 3.73* (2,57) |
| DEM | ITL | 0.19 (2,139) | 0.14 (2,18) | 0.17 (2,116) | 0.25 (2,138) | 0.36 (2,17) | 0.81 (2,15) |
| PTA | DEM | 3.39* (2,139) | 2.23 (2,54) | 0.41 (2,80) | 3.92* (2,138) | 2.47 (2,53) | 0.49 (2,79) |
| DEM | PTA | 0.42 (2,139) | 0.22 (2,18) | 0.22 (2,116) | 0.95 (2,138) | 0.47 (2,17) | 0.29 (2,115) |
| USD | DEM | 0.46 (2,139) | 0.12 (2,135) | 0.44 (2,99) | 1.30 (2,138) | 0.36 (2,34) | 1.35 (2,98) |
| DEM | USD | 4.75** (2,139) | 0.93 (2,18) | 3.12* (2,116) | 5.61** (2,138) | 0.89 (2,17) | 4.12* (2,115) |
The interest rates of the various currencies are represented by the following symbols: FRF, French franc; DEM, Deutsche mark; ITL, lira; PTA, peseta; USD, U.S. dollar.
The breaking points corresponding to changes in regime, as well as their degree of significance are presented in Table 2.
One and two asterisks indicate that X “Granger-causae” Y at levels of significance of 5 and 1 percent, respectively. Numbers in parentheses underlying F-statistics correspond to the degrees of freedom of the relevant test.
Quandt tests for possible changes in regime were carried out for the ERM currencies on Granger simple causality tests. The Quandt Log-likelihood functions are presented in Chart 1 under four panels and the corresponding breaking points determining changes in regime with associated Chow tests 7/ are reported in Table 2. The first panel concerns the causal relationship characterizing the influence of the French, Italian, and Spanish interest rates on the German rate. For each series, including the U.S. dollar, the maximum of the likelihood functions is reached in May 1981 and the Chow tests confirm the significance of this breaking point. This period marked the end of a phase of a relative weakness of the deutsche mark in the EMS and build-up of inflationary pressures in Germany that were met by important increases in the intervention rate of the Bundesbank at the outset of 1980 and 1981. The second, third, and fourth panels present the Log-likelihood functions of testing for possible change in the causal relationship accounting for the impact of the German interest rate on the French, Italian, and Spanish interest rates, respectively. A perusal of these charts reveals three breaking points: (i) August 1982 for the French rate, (ii) March 1986 for the Italian rate, and (iii) May 1984 for the Spanish rate. Except in the case of the Italian rate, the Chow test does not firmly support these results. Various factors can account for these changes in regime. Some of them are technical and related to changes in the choice of monetary instrument, while others are the expression of changes in the orientation of monetary policy. In the case of the French interest rate, the new regime started after the second realignment of the French franc in the EMS—the central rate of the French franc against deutsche mark was devalued by 10 percent in June 14, 1982—as the French government embarked on more restrictive financial measures and structural reforms in an attempt to counter inflationary pressures and restore the credibility of the stabilization of the French franc in the EMS. The strong increase in the Granger causality test between the two regimes confirms that the central element of the new policy was a greater coordination with German monetary policy. In the case of the lira, the existence of a breaking point in March 1986 appears to coincide with the relaxation of monetary policy as the lira declined in the ERM wide band from the strong position secured after the July 20, 1985 devaluation by 8 percent against all participating currencies. This appears to have been accompanied by a weakening of the influence of the German interest rate on the Italian rate as suggested by the Granger causality test. The participation of the lira in the ERM wide band—up to January 1990—and the implementation of capital controls seem to have been the main factors that allowed Italy to preserve a greater measure of monetary autonomy in the ERM than other members. In the case of the Spanish rate, the change in regime spotted in mid-1984 seems to correspond to the alteration in the implementation of monetary policy as the central bank switched from base money to interest rate control. This finding is consistent with the conclusions of Escriva and Santos (1991) who were unable to find a stable interest rate reaction function for the Bank of Spain before 1984. The Granger causality test does not provide significant evidence of a causal link between the German and Spanish interest rates, in particular after 1984. 8/ This should not come as a surprise in that the participation of Spain to the EC dates back to January 1986 and the entry of the peseta in the ERM wide band to June 1989. There is, however, recent evidence of the influence of other ERM interest rates, and in particular the German rate, on the Spanish interest rate in 1991–92 following the removal of remaining capital and credit controls. These developments are too recent to be taken into account by the Quandt likelihood test.
TEST OF CHANGE IN REGIME IN GRANGER CAUSALITY TESTS APPLIED INTEREST RATES OF ERM COUNTRIES, JAN 1980-DEC 1990
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
TEST OF CHANGE IN REGIME IN GRANGER CAUSALITY TESTS APPLIED INTEREST RATES OF ERM COUNTRIES, JAN 1980-DEC 1990
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
TEST OF CHANGE IN REGIME IN GRANGER CAUSALITY TESTS APPLIED INTEREST RATES OF ERM COUNTRIES, JAN 1980-DEC 1990
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
Determination of Change in Regime in Granger Causality Tests Using the Quandt Likelihood Test
The breaking points are determined by using the switching regression technique proposed by Quandt (1958), this technique is based on the calculation of a log-likelihood function assuming two regimes with different error variances and choosing the period for which this function is maximized.
One and two asterisks indicate that the existence of two regimes cannot be rejected at a level of significance of 5 and 1 percent, respectively.
Determination of Change in Regime in Granger Causality Tests Using the Quandt Likelihood Test
| ΔY | ΔX | Breaking Points 1/ | Chow Test 2/ |
|---|---|---|---|
| FRF | DEM | AUG 1982 | 1.89 |
| DEM | FRF | MAY 1981 | 5.20 (**) |
| LIT | DEM | MAR 1986 | 3.98 (**) |
| DEM | LIT | MAY 1981 | 2.84 (**) |
| PTA | DEM | MAY 1984 | 1.84 |
| DEM | PTA | MAY 1981 | 2.66 (*) |
| US | DEM | OCT 1982 | 0 |
| DEM | US | MAY 1981 | 2.3 (*) |
The breaking points are determined by using the switching regression technique proposed by Quandt (1958), this technique is based on the calculation of a log-likelihood function assuming two regimes with different error variances and choosing the period for which this function is maximized.
One and two asterisks indicate that the existence of two regimes cannot be rejected at a level of significance of 5 and 1 percent, respectively.
Determination of Change in Regime in Granger Causality Tests Using the Quandt Likelihood Test
| ΔY | ΔX | Breaking Points 1/ | Chow Test 2/ |
|---|---|---|---|
| FRF | DEM | AUG 1982 | 1.89 |
| DEM | FRF | MAY 1981 | 5.20 (**) |
| LIT | DEM | MAR 1986 | 3.98 (**) |
| DEM | LIT | MAY 1981 | 2.84 (**) |
| PTA | DEM | MAY 1984 | 1.84 |
| DEM | PTA | MAY 1981 | 2.66 (*) |
| US | DEM | OCT 1982 | 0 |
| DEM | US | MAY 1981 | 2.3 (*) |
The breaking points are determined by using the switching regression technique proposed by Quandt (1958), this technique is based on the calculation of a log-likelihood function assuming two regimes with different error variances and choosing the period for which this function is maximized.
One and two asterisks indicate that the existence of two regimes cannot be rejected at a level of significance of 5 and 1 percent, respectively.
All in all, Granger causality and Quandt tests implemented over the period July 1979 to June 1991 provide evidence of a significant influence of the German interest rate on French and Italian interest rates. After 1982, the tests point to the strengthening of the relationship between German and French interest rates associated with the development of a strong instantaneous causality between the two rates. Conversely, the fluctuations of the German interest rate appear to have been somewhat caused by those of the US interest rate after 1981.
2. Causal relationships in a co-integration framework
This section investigates the long-term causal relationship between the domestic interest rate of a given ERM country and the German interest rate corrected for inflation differentials using co-integration properties as suggested by Granger (1988). Under the German leadership hypothesis, ERM central banks, to establish the credibility of their anti-inflationary stance, would have been expected to secure inflation convergence with Germany by adjusting their domestic interest rates in relation both to variations of the German interest rate and to changes in expected inflation differentials with Germany. In the long run, such policy would translate into the cointegration of the interest rate of the ERM country with that of Germany corrected for inflation differentials. There are some indications that such a procedure was carried out on a regular basis, at least in one central bank, with inflation expectations based in part on recent trends and in part on forecasts issued every semester by international organizations.
To test for the presence of co-integration, a two-step procedure is adopted. First, to determine the existence of co-integration, ADF unit root tests are applied to each ERM interest rate and to its estimate, constructed with the German interest rate corrected for the relevant expected interest rate differentials. 9/ Second, after taking first differences, two regressions are estimated. The ERM interest rate is regressed on its own lagged values and on those of the German interest rate corrected for inflation differentials and vice versa. To the extent that the first step does not lead to a rejection of the assumption of co-integration, an error correction term is included in the regressions to account for the long-term component of the causal relationship. 10/ The error correction term is the co-integrating vector estimated by regressing the level of the actual interest rate on the estimated rate. The significance of a long-term causal linkage would be recognized if the ADF unit root tests are conclusive and if the regression coefficient on the error correction term is significantly different from zero.
Table 3 reports only the results of the first regression estimating the relationship between the French, Italian, and Spanish interest rates, and the German rate corrected for inflation differentials as the significance of the statistical tests on the opposite relationship were distinctly low. Results are presented both for the overall period and the second regime identified in Table 2 in the case of the deutsche mark. The T-student on the coefficient of the error correction term is generally significant at the 99 percent confidence level while the ADF tests point to the existence of co-integration with a level of significance rather low in the case of the Italian and Spanish rate for the overall period, and much stronger in the case of the French rate after June 1981. The latter result provides additional support for the previous evidence in favor of the German leadership thesis after 1981.
Granger Casuality Tests in a Co-integration Framework Applied to ERM Interest Rates
(Tests of Two Regimes Over the period july 1979 to June 1991)
Y is the actual interest rate and x the estimated interest rate determined by adding up to the German interest rate, the contemporaneous 12-month inflation differentials between the relevant country and Germany.
One and two asterisks in columns (2), (3), (8) and (9) indicate significance levels of 5 and 1 percent, respectively.
ADF critical values for unit root tests calculated in Engle and Granger (1987) for 100 obversatios are 3.77, 3.17 and 2.84 at a significance level of 1 percent, 5 percent, and 10 percent, respectively, Co-integration at 10 and 5 percent levels of significance is indicated by one and two asterisks, respectively, in columns (6) and (12).
Granger Casuality Tests in a Co-integration Framework Applied to ERM Interest Rates
(Tests of Two Regimes Over the period july 1979 to June 1991)
| OVERALL PERIOD 2/ | REGIME 2 (June 1981 to June 1991) 2/ | ||||||||||||
|---|---|---|---|---|---|---|---|---|---|---|---|---|---|
| ADF Unit Root Test 3/ | ADF Unit Root Test 3/ | ||||||||||||
| ΔY | ΔX | Theil Adjusted R2 | F-Statistic | t-student on z | y | x | Theil Adjusted R2 | F-Statistic | t-student on z | y | x | ||
| (1) | (2) | (3) | (4) | (5) | (6) | (7) | (8) | (9) | (10) | (11) | (12) | ||
| FRF | Estimated FRF | 0.285 | 9.14(**) | 3.52(**) | 1.66 | 1.06 | 2.53 | 0.242 | 6.31(**) | 2.36(*) | 1.99 | 2.85 | 3.47(**) |
| ITL | Eetimated ITL | 0.219 | 6.75(**) | 3.16(**) | 1.47 | 0.77 | 2.90(*) | 0.209 | 5.40(**) | 2.85(**) | 1.16 | 4.84 | 2.29 |
| PTA | Estimated PTA | 0.285 | 9.17(**) | 3.82(**) | 2.97 | 1.49 | 3.01(*) | 0.253 | 6.64(**) | 3.07(**) | 2.71 | 2.84 | 2.60 |
Y is the actual interest rate and x the estimated interest rate determined by adding up to the German interest rate, the contemporaneous 12-month inflation differentials between the relevant country and Germany.
One and two asterisks in columns (2), (3), (8) and (9) indicate significance levels of 5 and 1 percent, respectively.
ADF critical values for unit root tests calculated in Engle and Granger (1987) for 100 obversatios are 3.77, 3.17 and 2.84 at a significance level of 1 percent, 5 percent, and 10 percent, respectively, Co-integration at 10 and 5 percent levels of significance is indicated by one and two asterisks, respectively, in columns (6) and (12).
Granger Casuality Tests in a Co-integration Framework Applied to ERM Interest Rates
(Tests of Two Regimes Over the period july 1979 to June 1991)
| OVERALL PERIOD 2/ | REGIME 2 (June 1981 to June 1991) 2/ | ||||||||||||
|---|---|---|---|---|---|---|---|---|---|---|---|---|---|
| ADF Unit Root Test 3/ | ADF Unit Root Test 3/ | ||||||||||||
| ΔY | ΔX | Theil Adjusted R2 | F-Statistic | t-student on z | y | x | Theil Adjusted R2 | F-Statistic | t-student on z | y | x | ||
| (1) | (2) | (3) | (4) | (5) | (6) | (7) | (8) | (9) | (10) | (11) | (12) | ||
| FRF | Estimated FRF | 0.285 | 9.14(**) | 3.52(**) | 1.66 | 1.06 | 2.53 | 0.242 | 6.31(**) | 2.36(*) | 1.99 | 2.85 | 3.47(**) |
| ITL | Eetimated ITL | 0.219 | 6.75(**) | 3.16(**) | 1.47 | 0.77 | 2.90(*) | 0.209 | 5.40(**) | 2.85(**) | 1.16 | 4.84 | 2.29 |
| PTA | Estimated PTA | 0.285 | 9.17(**) | 3.82(**) | 2.97 | 1.49 | 3.01(*) | 0.253 | 6.64(**) | 3.07(**) | 2.71 | 2.84 | 2.60 |
Y is the actual interest rate and x the estimated interest rate determined by adding up to the German interest rate, the contemporaneous 12-month inflation differentials between the relevant country and Germany.
One and two asterisks in columns (2), (3), (8) and (9) indicate significance levels of 5 and 1 percent, respectively.
ADF critical values for unit root tests calculated in Engle and Granger (1987) for 100 obversatios are 3.77, 3.17 and 2.84 at a significance level of 1 percent, 5 percent, and 10 percent, respectively, Co-integration at 10 and 5 percent levels of significance is indicated by one and two asterisks, respectively, in columns (6) and (12).
3. Reaction functions
A reaction function was estimated for each currency, except the peseta. To take into account the short- and long-term relationships observed in the causality tests, the reaction function was estimated as an error correction model including both the levels, lagged by one period, and changes of the explanatory variables identified in the previous tests. Other relevant explanatory variables were included in the reaction function on the basis of further empirical evidence and results of previous studies (Galy (1985), Artus and alii (1987), Artus and alii (1991)). The following explanatory variables were added: (i) the ratio of foreign reserves to imports; (ii) the deviations of the exchange rates of the French franc and Italian lira from their ERM central rates against the deutsche mark; (iii) the exchange rate of the deutsche mark against the U.S. dollar.
Table 4 presents the reaction functions while the long-term elasticities implied by the correction terms are reported in Table 5. The Quandt test was also applied to the reaction functions and the conclusions provided on the breaking points confirmed broadly the results provided by the causality tests, except for the Italian rate for which the maximum of the likelihood function is now reached in June 1981. This apparent change in regime follows the first devaluation of the lira in the ERM on March 23, 1981 (Chart 3, Table 6). It seems that after the two first years of the ERM, the Italian authorities paid greater attention to the constraints imposed on domestic monetary policy through participation in the ERM. This is suggested by the greater significance acquired by the German interest rate in the reaction function after June 1981.
ACTUAL AND ESTIMATTED MONEY MARKET INTEREST RATES IMPLIED BY THE PPP AND UIP PARADIGMS, JAN 1980-AUG 1991
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
ACTUAL AND ESTIMATTED MONEY MARKET INTEREST RATES IMPLIED BY THE PPP AND UIP PARADIGMS, JAN 1980-AUG 1991
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
ACTUAL AND ESTIMATTED MONEY MARKET INTEREST RATES IMPLIED BY THE PPP AND UIP PARADIGMS, JAN 1980-AUG 1991
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
TESTS OF CHANGE IN REGIME IN THE INTEREST RATE REACTION FUNCTION OF ERM COUNTRIES, JUN 1980-DEC 1990
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
TESTS OF CHANGE IN REGIME IN THE INTEREST RATE REACTION FUNCTION OF ERM COUNTRIES, JUN 1980-DEC 1990
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
TESTS OF CHANGE IN REGIME IN THE INTEREST RATE REACTION FUNCTION OF ERM COUNTRIES, JUN 1980-DEC 1990
Citation: IMF Working Papers 1992, 087; 10.5089/9781451850932.001.A001
Estimation of the Reaction Functions for Short-term ERM Interests Rates 1/
The breaking points and their degree of significance are presented in Table 6.
All variables are expressed in natural logarithms, except interest rates and inflation rates. The inflation differential is the 12-month inflation differential between France (Italy) and Germany for the French (Italian) reaction function, and between Germany and France for the German reaction function. The deviation from central rates is the ratio of the exchange rate of the French franc in terms of one unit of deutsche mark to its ERM central rates. The exchange rate of the deutsche mark is expressed per unit of U.S. dollar. Coefficients in parentheses are t-ratios. One and two asterisks point to level of significance of 5 and 1 percent, respectively. F, R2, DW, and SEE denote the F, test, the coefficient of determination, the Durbin-Matson statistic and the standard error.
The β’s coefficients are long-term elasticities. They are estimated by dividing the coefficient estimated on the variable in level (part II of the table) by the error correction term of the relevant endogenous interest rate. The long-term elasticities are presented in Table 5.
Estimation of the Reaction Functions for Short-term ERM Interests Rates 1/
| Change in French interest rate (Δi) 1/ | Change in German interest rate (Δi) 1/ | Change in Italian interest rate (Δi) | |||||||
|---|---|---|---|---|---|---|---|---|---|
Exogenous Variables / Endogenous Variables | Regime 1 7–1979 7–1981 | Regime 2 8–1981 10–1991 | Regime 1 7–1979 3–1981 | Regime 2 4–1981 10–1991 | Regime 1 7–1979 5–1981 | Regime 2 6–1981 10–1991 | |||
| I. First differences (in t-1)2/ | |||||||||
| Δi* - italian interest rate | — | — | 0.13 (0.53) | 0.11* (2.39) | 0.09 (0.44) | 0.41** (4.89) | (in t-1) | ||
| Δi* - French interest rate | — | — | 0.51 (1.43) | 0.17** (3.41) | — | — | |||
| Δi* - German interest rate | 0.23 (1.38) | 0.23* (2.47) | — | — | 0.12 (0.72) | 0.27 (1.94) | |||
| Δ(p-p*) - Inflation differential | 0.01 (0.04) | 0.24** (2.99) | 0.08 (0.21) | 0.05 (0.74) | 0.13 (0.73) | 0.11 (1.11) | |||
| Δ(s-Ṡ) - Deviation from central rates | -0.09 (0.52) | 0.11** (3.33) | — | — | — | ||||
| Δ(sus) - U.S. dollar (in t-1) | — | — | -0.057 (0.85) | (0.023** (3.49) | — | — | |||
| Δr - Foreign reserves/import ratio | -0.04** (4.31) | -0.005** (3.81) | -0.019 (0.48) | -0.017** (3.55) | -0.02 (1.32) | -0.013* | |||
| II. Error correction term: variables in level in t-13/ | |||||||||
| i, i* -Italian interest rate | — | — | — | — | 0.53**(3.55) | -0.06* | |||
| i, i* - French interest rate | -0.26* (2.13) | -0.15** (3.65) | 0.41 (1.71) | 0.06* (2.47) | — | — | |||
| i, i* - German interest rate | 0.21 (1.75) | 0.08* (2.57) | -0.36 (1.16) | -0.07** (3.14) | 0.42** (3.12) | 0.04 (2.00) | |||
| (p-p*) - Inflation differential | -0.05 (0.16) | 0.12** (3.73) | 0.37 (1.50) | 0.07** (4.40) | 0.44** (2.94) | 0.08* (2.02) | |||
| (s-Ṡ) - Deviation from central rates | 0.21 (0.98) | 0.11** (3.49) | — | — | — | — | |||
| sus - U.S. dollar | — | — | 0.11* (2.47) | 0.003 (1.95) | — | — | |||
| r - Foreign reserves/import ratio | -0.003 (1.19) | -0.001** (3.36) | -0.048* (2.32) | -0.002 (1.77) | -0.01 (1.51) | -0.001 (0.62) | |||
| III. Statistical parameters | |||||||||
| - F-statistic | 7.10 | 10.35 | 2.99 | 7.91 | 4.39 | 5.287 | |||
| - R2 (adjusted) | 0.66 | 0.39 | 0.42 | 0.35 | 0.54 | 0.21 | |||
| - DW (or H statistic for Italian lira) | 1.95 | 1.80 | 1.73 | 1.83 | -5.70 | 0.15 | |||
| - SEE | 0.006 | 0.003 | 0.003 | 0.003 | 0.006 | 0.004 | |||
| - Number of observations | 25 | 123 | 21 | 127 | 23 | 148 | |||
The breaking points and their degree of significance are presented in Table 6.
All variables are expressed in natural logarithms, except interest rates and inflation rates. The inflation differential is the 12-month inflation differential between France (Italy) and Germany for the French (Italian) reaction function, and between Germany and France for the German reaction function. The deviation from central rates is the ratio of the exchange rate of the French franc in terms of one unit of deutsche mark to its ERM central rates. The exchange rate of the deutsche mark is expressed per unit of U.S. dollar. Coefficients in parentheses are t-ratios. One and two asterisks point to level of significance of 5 and 1 percent, respectively. F, R2, DW, and SEE denote the F, test, the coefficient of determination, the Durbin-Matson statistic and the standard error.
The β’s coefficients are long-term elasticities. They are estimated by dividing the coefficient estimated on the variable in level (part II of the table) by the error correction term of the relevant endogenous interest rate. The long-term elasticities are presented in Table 5.
Estimation of the Reaction Functions for Short-term ERM Interests Rates 1/
| Change in French interest rate (Δi) 1/ | Change in German interest rate (Δi) 1/ | Change in Italian interest rate (Δi) | |||||||
|---|---|---|---|---|---|---|---|---|---|
Exogenous Variables / Endogenous Variables | Regime 1 7–1979 7–1981 | Regime 2 8–1981 10–1991 | Regime 1 7–1979 3–1981 | Regime 2 4–1981 10–1991 | Regime 1 7–1979 5–1981 | Regime 2 6–1981 10–1991 | |||
| I. First differences (in t-1)2/ | |||||||||
| Δi* - italian interest rate | — | — | 0.13 (0.53) | 0.11* (2.39) | 0.09 (0.44) | 0.41** (4.89) | (in t-1) | ||
| Δi* - French interest rate | — | — | 0.51 (1.43) | 0.17** (3.41) | — | — | |||
| Δi* - German interest rate | 0.23 (1.38) | 0.23* (2.47) | — | — | 0.12 (0.72) | 0.27 (1.94) | |||
| Δ(p-p*) - Inflation differential | 0.01 (0.04) | 0.24** (2.99) | 0.08 (0.21) | 0.05 (0.74) | 0.13 (0.73) | 0.11 (1.11) | |||
| Δ(s-Ṡ) - Deviation from central rates | -0.09 (0.52) | 0.11** (3.33) | — | — | — | ||||
| Δ(sus) - U.S. dollar (in t-1) | — | — | -0.057 (0.85) | (0.023** (3.49) | — | — | |||
| Δr - Foreign reserves/import ratio | -0.04** (4.31) | -0.005** (3.81) | -0.019 (0.48) | -0.017** (3.55) | -0.02 (1.32) | -0.013* | |||
| II. Error correction term: variables in level in t-13/ | |||||||||
| i, i* -Italian interest rate | — | — | — | — | 0.53**(3.55) | -0.06* | |||
| i, i* - French interest rate | -0.26* (2.13) | -0.15** (3.65) | 0.41 (1.71) | 0.06* (2.47) | — | — | |||
| i, i* - German interest rate | 0.21 (1.75) | 0.08* (2.57) | -0.36 (1.16) | -0.07** (3.14) | 0.42** (3.12) | 0.04 (2.00) | |||
| (p-p*) - Inflation differential | -0.05 (0.16) | 0.12** (3.73) | 0.37 (1.50) | 0.07** (4.40) | 0.44** (2.94) | 0.08* (2.02) | |||
| (s-Ṡ) - Deviation from central rates | 0.21 (0.98) | 0.11** (3.49) | — | — | — | — | |||
| sus - U.S. dollar | — | — | 0.11* (2.47) | 0.003 (1.95) | — | — | |||
| r - Foreign reserves/import ratio | -0.003 (1.19) | -0.001** (3.36) | -0.048* (2.32) | -0.002 (1.77) | -0.01 (1.51) | -0.001 (0.62) | |||
| III. Statistical parameters | |||||||||
| - F-statistic | 7.10 | 10.35 | 2.99 | 7.91 | 4.39 | 5.287 | |||
| - R2 (adjusted) | 0.66 | 0.39 | 0.42 | 0.35 | 0.54 | 0.21 | |||
| - DW (or H statistic for Italian lira) | 1.95 | 1.80 | 1.73 | 1.83 | -5.70 | 0.15 | |||
| - SEE | 0.006 | 0.003 | 0.003 | 0.003 | 0.006 | 0.004 | |||
| - Number of observations | 25 | 123 | 21 | 127 | 23 | 148 | |||
The breaking points and their degree of significance are presented in Table 6.
All variables are expressed in natural logarithms, except interest rates and inflation rates. The inflation differential is the 12-month inflation differential between France (Italy) and Germany for the French (Italian) reaction function, and between Germany and France for the German reaction function. The deviation from central rates is the ratio of the exchange rate of the French franc in terms of one unit of deutsche mark to its ERM central rates. The exchange rate of the deutsche mark is expressed per unit of U.S. dollar. Coefficients in parentheses are t-ratios. One and two asterisks point to level of significance of 5 and 1 percent, respectively. F, R2, DW, and SEE denote the F, test, the coefficient of determination, the Durbin-Matson statistic and the standard error.
The β’s coefficients are long-term elasticities. They are estimated by dividing the coefficient estimated on the variable in level (part II of the table) by the error correction term of the relevant endogenous interest rate. The long-term elasticities are presented in Table 5.
Long-Term Elasticities in the Reaction Functions with Error Correction Terms 1/
Long-term elasticities are determined on the basis of the error correction terms in Table 4.
Long-Term Elasticities in the Reaction Functions with Error Correction Terms 1/
| French interest rate | German interest rate | Italian interest rate | ||
|---|---|---|---|---|
| 8/1981 to 10/1991 | 4/1981 to 10/1991 | 6/1981 to 10/1991 | ||
| Explanatory variables | ||||
| - French interest rate | 0.86 | |||
| - German interest rate | 0.53 | 0.67 | ||
| - Inflation differential | 0.80 | 1.00 | 1.33 | |
| - Deviation from central rates | 0.73 | |||
| - U.S. dollar | 0.04 | |||
| - Foreign reserves/import ratio | -0.007 | -0.03 | NS | |
Long-term elasticities are determined on the basis of the error correction terms in Table 4.
Long-Term Elasticities in the Reaction Functions with Error Correction Terms 1/
| French interest rate | German interest rate | Italian interest rate | ||
|---|---|---|---|---|
| 8/1981 to 10/1991 | 4/1981 to 10/1991 | 6/1981 to 10/1991 | ||
| Explanatory variables | ||||
| - French interest rate | 0.86 | |||
| - German interest rate | 0.53 | 0.67 | ||
| - Inflation differential | 0.80 | 1.00 | 1.33 | |
| - Deviation from central rates | 0.73 | |||
| - U.S. dollar | 0.04 | |||
| - Foreign reserves/import ratio | -0.007 | -0.03 | NS | |
Long-term elasticities are determined on the basis of the error correction terms in Table 4.
Determination of Change in Regime in the Interest Rate Reaction Functions using the Quandt Likelihood Test
See Table 2.
See Table 2.
The perusal of the reaction functions estimated over the two regimes suggests the following observations:
- In the French interest rate equation, only foreign reserves stand out as a strong explanatory factor during the first regime. By contrast, after July 1981, all exogenous variables, including the error correction terms become highly significant. The German interest rate and the inflation differentials have a notable influence both in the short-term dynamics and in the long-term components of the reaction function. The long-term elasticities emerging from the error correction terms (Table 6) point to the important role played by the inflation differentials with Germany and to a lesser degree by the German interest rate in the long-run orientation of the French monetary policy. Also, the intensity of the reaction of the French central bank increased significantly over the second regime with respect to deviations of the French franc from its central rate against the deutsche mark but weakened with respect to changes in foreign reserves. This suggests that, during this period, the authorities relied more on monetary policy than on exchange rate policy to stabilize the exchange rate. This outcome is consistent with the decline of the effectiveness of sterilized foreign exchange intervention that was prompted by greater capital mobility and substitutability as capital controls were progressively phased out. While the coefficient of the German interest rate remained stable over the two regimes, suggesting the absence of changes in the behavior of the central bank in this regard, the fact that the coefficient of the deviations of the French franc from its deutsche mark central rate increased strikingly during the second regime, could be interpreted as a relative tightening of the constraint imposed by German monetary policy on French monetary policy.
- In the German interest rate equation, as in the French one, most of the explanatory variables are hardly significant or possess signs that disagree with a priori expectations during the first regime. Over the second period, most of the coefficients turn out to be highly significant. The short-term dynamics of the German interest rate are explained by the reaction of the Bundesbank to changes in: (i) the French and Italian interest rates; (ii) the exchange rate of the deutsche mark against the U.S. dollar; (iii) and foreign reserves. The long-term trend of the German interest rate is significantly influenced by the French interest rate and inflation differentials, mirroring in part the results obtained in the French interest rate reaction function.
- In the Italian reaction function, short-term explanatory variables are not significant during the first regime. Over the second regime, only the lagged changes of the Italian interest rate and foreign reserves appear to drive the short-term dynamics of the Italian rate. The level of significance of the German interest rate, while improving substantially compared to the first regime, remains weak and deviations from central rates are not significant at all. Also, the error correction elements exhibit a modest degree of significance. These results seem to illustrate the relative independence enjoyed by the Italian central bank during the 1980s, thanks to capital controls and to the relatively limited constraint imposed by the participation of the lira in the ERM broad band.
V. Conclusion
Econometric evidence presented in this paper provides some support to the German leadership thesis but points also to an increased interdependence between French and German monetary policies after 1981–82 while results in the Italian and Spanish case suggest the persistence of a significant monetary autonomy. Changes in regime appear to be concentrated in the first years of the ERM and are related to technical changes in the implementation of monetary policy, and to the introduction of adjustment programs to strengthen the credibility of the exchange rate commitment, generally following a major realignment. There is no doubt that the total removal of capital controls in 1990–91 has reduced further monetary autonomy among ERM participants, including Germany, requiring a rapid implementation of the concerted approach proposed in the Maastricht Agreement. However, the consecutive possible changes in ERM central banks’ behavior are too recent to be identified by the tests presented in this paper.
References
Artus, P., Y. Barroux, and J. Pecha, “Objectif Quantitatif de Croissance de la Masse Monetaire et Fonction de Reaction des Autorites,” Cahiers Economiques et Monetaires, Banque de France, No. 27, pp. 49–58, 1987.
Artus, P., Y. Barroux, and J. Pecha, S. Avouyi-Dovi, E. Bleuze, and F. Lecointe, “Transmission of U.S Monetary Policy to Europe and Asymmetry in the European Monetary System,” European Economic Review, No. 35, pp. 1369–1384, 1991.
De Grauwe, P., “Is the European Monetary System a DM-zone?” CEPR discussion paper, No. 297, March 1989.
Engle, R.F., and C.W.J. Granger, “Co-integration and Error Correction: Representation, Estimation, and Testing,” Econometrica, Vol. 55, No. 2, pp. 251–276, March 1987.
Escriva, J.L., and R. Santos, “Un Estudio del Cambio de Regimen en la Variable Instrumental del Control Monetario en España,” Documento de Trabajo No. 9111, Servicio de Estudios, Banco de España, 1991.
Fisher, S., “International Macroeconomic Policy Coordination,” NBER Working Paper No. 2244, 1987.
Galy, M., “Evaluation du Processus d’ Integration Monetaire au Sein du Systeme Monetaire Europeen,” Cahiers Economiques et Monetaires, Banque de France, No. 20, pp. 59–88, 1985.
Gardner, E., and W. Perraudin, “Asymmetry within the ERM: A Case Study of French and German Interest Rates Since Basel-Nyborg,” IMF Working Paper forthcoming, 1992.
Giavazzi, F., and A. Giovannini, “Models of the EMS: Is Europe a Greater Deutschemark Area?” in R. C. Bryant, and, R. Portes, editors, Global Macroeconomics, New-York, St. Martin’s Press, 1987.
Granger, C.W.J., “Investigating Causal Relations by Econometric Models and Cross-Spectral Methods,” Econometrica, Vol. 37, No. 3, pp. 425–438, July 1969.
Granger, C.W.J., “Some Recent Developments in a Concept of Causality”, Journal of Econometrics, No. 39, pp. 199–211, 1988.
Melitz, J., “The Prospect of a Depreciating Dollar and Possible Tension Inside the EMS,” CEPR Discussions Paper No. 97, March 1986.
Quandt, R.E., “The Estimation of the Parameters of a Linear Regression System Obeying Two Separate Regimesm,” Journal of the American Statistical Association, pp. 873–880, 1958.
Sargent, T. J., “A Classical Macroeconomic Model for the United States,” Journal of Political Economy, Vol. 84, No. 2, pp. 207–237, 1976.
Von Hagen, J., and M. Fratianni, “The European Monetary System Ten Years After,” in C. Plosser, and A. H. Meltzer, editors, Carnegie Rochester Conference Series on Public Policy, 32, 1990 a.
Von Hagen, J., and M. Fratianni, “German Dominance in the EMS: Evidence from Interest Rates,” Journal of International Money and Finance, 9, pp. 358–375, 1990 b.
The author thanks Thomas Krueger and members of the European I Department Luncheon Seminar for helpful suggestions.
It is true that intra-marginal interventions were more frequent than interventions at the margins. However, the size of the latter were by far more substantial.
Interest rates originate from the IFS database. They are defined as follows: the three-month interbank rate for France, Germany, and Italy; the intervention rate at the central bank’s auction for Spain; the federal funds rate for the United States.
It is true, however, that co-integration of a given ERM interest rate with the German interest rate corrected by inflation differentials—with a co-integrating constant equal to 1—would also apply if the Uncovered Interest Parity and the Purchasing Power Parity paradigms hold (Chart 2).
For each regression, the following Log-likelihood function was estimated:
where σ1 and σ2 are the standard errors of the regression under the first and second regimes; t is the observation corresponding to a change in regimes, and T is the total number of observations. The Log-likelihood function is estimated for each t. The period
Unit root tests provided evidence of non-stationarity for all exogenous and endogenous variables used in this paper. To avoid spurious correlation in the estimation of causality tests and reaction functions, all variables were N-differenced so as to render them stationary. Interest rate series are first-differenced.
Note that the Chow test is not totally consistent with the Quandt Log-likelihood test as the Chow test relies on the assumption of equal error variances over the two regimes, which is not the case for the Quandt test.
The fact that there is no causality on the basis of short-term interest rates between the German rate and other rates does not exclude any stronger linkages for other maturities. It is assumed, here, that such linkages are more likely to be the results of the increase in international capital market integration rather than induced by the behavior of central banks.
Assuming perfect foresight, expected inflation is equal to ex-post inflation.
Engle and Granger (1987) proved that cointegrated series can be represented by error correction models. In the two variable case, the generating dynamics can be expressed as:
where
