Abstract
This paper makes three contributions. First we present a technique by which the monetary transmission mechanism of Germany, France, the UK and the Eurozone can be decomposed into its component cycles, compared across economies and across time. As a result, we found that the individual data generating processes have varied over time. Second we show that Germany has now converged on the rest of Europe and not vice versa, although Germany had dominated monetary policy making in Europe for many years. Third, we show that the UK as an outsider has behaved like a peripheral EMU country, even when EMU was not in place. In other words, the transmission mechanisms of Germany and the UK were fundamentally different. Hence, when that German monetary policy dominated Europe in a way that was not in line with the rest of Europe, never mind the UK, it is no surprise that the UK eventually left the ERM (1992). The current financial crisis may enforce the trend of convergence of the transmission mechanism. But there have been signs of a divergence between core and periphery, to some extent involving the UK, so this general convergence, as opposed to tighter convergence in the core, may not last.
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Notes
BIS (1995), Dale and Haldane (1995), Ramaswamy and Sloek (1998), Hughes Hallett and Piscitelli (1999, 2002). Asymmetries in monetary transmissions can arise from variations in house ownership and household debt or mortgages (MacLennan et al. 2000); from variations in the financing of firms and hence the industrial structures in different economies (Kashyup and Stein 1997; Carlino and DeFina 1999); or from variations in the size of firms (industrial concentration), financial structures and the legal system (Cecchetti 1999) – in short, from both the borrowing and the lending side of monetary transactions. Indeed, MacLennan et al. (2000) argue that such “indirect” impacts of monetary asymmetries are likely to prove more difficult for policy to absorb that a lack of synchronisation in business cycles.
Obviously, using the entire sample implies that we neglect possible structural breaks. The initial estimates may be biased therefore. The Kalman filter will then correct for this since, as Wells (1996) points out, the Kalman filter will converge to the true parameter value independently of the initial value. But choosing initial values which are already “close” to the true value accelerates convergence. Hence we employ an OLS estimate to start. And the start values have no effect on the parameter estimates by the time we get to 2008. Our results are robust.
Notice that all our tests of significance, and significant differences in parameters, are being conducted in the time domain, before transferring to the frequency domain, because no statistical tests exist for calculated spectra (the transformations being nonlinear and involving complex arithmetic). Stability tests are important here because our spectra are sensitive to changes in the underlying parameters (see Section 3).
The fluctuations test works as follows: one parameter value is taken as the reference value, e.g. the last value of the sample. All other observations are now tested whether they significantly differ from that value. In order to do so, Ploberger et al. (1989) have provided critical values that we have used. If the test value is above the critical value then we have a structural break, i.e. the parameter value differs significantly from the reference value.
This is not a complete picture of the transmission mechanism of course, but it does represent a necessary condition for convergence. Had we also been interested in similarities in the incidence or timing of the impacts of monetary policy changes, then we would have to compare phase shifts across countries too.
Malta and Slovenia were omitted in order to avoid another structural break.
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Appendices
Appendix 1: Short time fourier transform
Consider a signal s(τ) and a real, even window w(τ), whose Fourier transforms are S(f) and W(f) respectively. To obtain a localised spectrum s(τ) at time τ = t, we multiply the signal by the window w(τ) centred at time τ = t. We obtain
We then calculate the Fourier transform w.r.t. τ which yields
\( {\text{F}}_{\text{s}}^w \left( {{\text{t}},{\text{f}}} \right) \) is the STFT. It transforms the signal into the frequency domain across time. It is therefore a function of both. Using a bilinear kernel and a Gabor transform (the time series is stationary, but may contain parameter changes), Boashash and Reilly (1992) show that the STFT can always be expressed as a time-varying discrete fast-Fourier transform calculated for each point in time. That has the convenient property that the “traditional” formulae for the coherence or the gain are still valid, but have to be recalculated at each point in time. The time-varying spectrum of the growth rate series can therefore be calculated as (see also: Lin 1997):
where ω is angular frequency and j is a complex number and α are the estimated coefficients. The main advantage of this method is that, at any point in time, a power spectrum can be calculated instantaneously from the updated parameters of the model (see also Lin 1997). Similarly, the power spectrum for any particular time interval can be calculated by averaging the filter parameters over that interval.
Appendix 2: Statistical results
Note: For reasons of space, the results quoted in the tables describe the final regression done and its diagnostic tests. But the figures which follow display the period by period spectral results.
Table 1 Regression results: German growth rate and Euro interest rates
VAR/System—Estimation by Kalman Filter | |||
Dependent Variable | DLGERGDP | Quarterly Data From | 1977:01 To 2007:02 |
Usable Observations | 122 | Std Error of Dependent Variable | 7.344034908 |
R2 | 0.98837 | Standard Error of Estimate | 4.953864717 |
Mean of Dependent Variable | −0.037649496 | Sum of Squared Residuals | 2871.2707493 |
Akaike (AIC) Criterion | 0.68802 | Ljung-Box Test: Q*(21) | 33.0404 |
Variable | Coeff | Std Error | T-Stat |
Constant | 0.92915083 | 0.069017425053 | 13.46255430988 |
DLGERGDP{1} | −0.83350102 | 0.862330010393 | −0.966568496268 |
GERINT | −0.51877805 | 0.180640176422 | −2.87188631575 |
GERINT{1} | 0.12575100 | 0.037446995505 | 3.358106480903 |
Table 2 Regression results: the French growth rate and Euro interest rates
VAR/System—Estimation by Kalman Filter | |||
Dependent Variable | DLFRGDP | Quarterly Data From | 1972:01 To 2008:01 |
Usable Observations | 144 | Std Error of Dependent Variable | 1.8633178724 |
R2 | 0.99882 | Standard Error of Estimate | 5.7263454851 |
Mean of Dependent Variable | 0.0083842572 | Sum of Squared Residuals | 4623.5355987 |
Akaike (AIC) Criterion | 0.70189 | Ljung-Box Test: Q*(26) | 35.7327 |
Variable | Coeff | Std Error | T-Stat |
Constant | 0.14457851 | 0.030366950564 | 4.7610481630 |
DLFRGDP{2} | 3.03444661 | 2.248747036208 | 1.349394378080 |
FRINT | 0.18960241 | 0.012935754672 | 14.65723611884 |
FRINT{1} | −0.217076786 | 0.072575370699 | −2.99105302958 |
Table 3 Regression results: the EMU growth rate and Euro interest rates
VAR/System—Estimation by Kalman Filter | |||
Dependent Variable | DLEMUGDP | Quarterly Data From | 1970:01 To 2008:01 |
Usable Observations | 145 | Std Error of Dependent Variable | 1.8916279266 |
R2 | 0.40258 | Standard Error of Estimate | 2.6225880721 |
Mean of Dependent Variable | 0.0023331688 | Sum of Squared Residuals | 962.91554742 |
Akaike (AIC) Criterion | 0.01512 | Ljung-Box Test: Q*(24) | 22.6655 |
Variable | Coeff | Std Error | T-Stat |
Constant | −0.09454859 | 0.038465000620 | −2.45804203950 |
DLEMUGDP{3} | −0.76789776 | 2.549985465622 | −0.301138094484 |
EMUINT | 0.12742306 | 0.076250929192 | 1.671101732675 |
EMUINT{1} | −0.26594720 | 0.035038487742 | −7.59014483447 |
Table 4 Regression results: the UK growth rate and Euro interest rates
VAR/System — Estimation by Kalman Filter | |||
Dependent Variable | DLUKGDP | Quarterly Data From | 1966:01 To 2008:01 |
Usable Observations | 169 | Std Error of Dependent Variable | 4.3055685240 |
R2 | 0.98995 | Standard Error of Estimate | 4.9772938011 |
Mean of Dependent Variable | 0.0023204919 | Sum of Squared Residuals | 4038.0729340 |
Akaike (AIC) Criterion | 0.20073 | Ljung-Box Test: Q*(26) | 20.2142 |
Variable | Coeff | Std Error | T-Stat |
Constant | 0.685210687 | 1.171592652093 | 0.584854032314 |
DLUKGDP{2} | 0.817373815 | 2.650584706442 | 0.308374907941 |
DLUKGDP{8} | −0.120035372 | 0.023446142261 | −5.11962142703 |
UKINT{2} | 0.163644836 | 0.033876378444 | 4.830647294398 |
UKINT{5} | −0.310839329 | 0.055562194434 | −5.59443938632 |
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Hughes Hallett, A., Richter, C. Has there been any structural convergence in the transmission of European monetary policies?. Int Econ Econ Policy 6, 85–101 (2009). https://doi.org/10.1007/s10368-009-0132-5
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DOI: https://doi.org/10.1007/s10368-009-0132-5