Abstract
We investigate the impact of the financial cycle on fiscal policy by estimating fiscal multipliers for different types of government spending that are contingent on two states determined by the financial cycle. To obtain our estimates, we extend the threshold VAR method from a single country to a panel of high-income countries. Our results indicate that the multiplier for government investment is affected by the state of the financial cycle: while the initial impact is positive for both states, in an upturn it turns negative, while in a downturn it remains positive. In the case of government consumption, the multiplier does not seem to significantly depend on the financial cycle. However, jointly conditioning on the financial and business cycles produces multipliers of government consumption which vary over the states of both cycles. This contrasts with the results for government investment which are left essentially unchanged by the joint conditioning on the four states defined by both cycles.
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Notes
The term “tight” for the credit regime suggests that the mechanism is supply driven, whereas it is also possible that the low level of credit is at least partially due to low demand.
Our code is based on the univariate TVAR estimation program by Gabriel Bruneau, freely available from his homepage.
We experimented with regime-dependent averages, this being the most flexible specification. This did not influence our main results in any noteworthy fashion. Results are available upon request.
Replacing the real interest rate by the nominal rate, to allow for the strict monetary policy reaction, does not alter the results.
Australia, Belgium, Canada, Denmark, Finland, France, Germany, Ireland, Italy, Japan, the Netherlands, Norway, South Korea, Spain, Sweden, Switzerland, UK, USA.
These data are based on the detrended credit-to-GDP gap using the HP-filter with \(\lambda _{\textrm{HP}}= 4\)00,000. We obtained these detrended series from the BIS (Drehmann et al. 2016). The relatively large value of \(\lambda _{\textrm{HP}}\) they use to detrend the credit gap only removes the long-term trend, leaving the longer frequency cycle associated with the financial cycle in their series. This approach, however, also allows the higher frequencies associated with the business cycle to pass through the filter and remain in their financial cycle measure. To obtain a cleaner series containing only the longer financial cycle frequencies, we HP-filter the BIS’s series again with \(\lambda _{\textrm{HP}}=1600\) and use the resulting HP-filter trend component to obtain a smoother financial cycle. In this way we remove the higher business cycle frequencies from the BIS’s series.
The Schüler cycles we received from the authors are smoothed using a six-quarter one-sided moving average based on the weights of a Bartlett window. We demean the Schüler cycles to ensure that they fluctuate around zero, instead of 0.5 as in Schüler et al. (2020). Their approach, the power cohesion method, maps the growth rates of financial variables (such as credit volumes and house prices) into an empirical cumulative distribution function between zero and one. Therefore, these series do not exhibit long-term trends.
The BIS kindly provided us with their financial cycle estimates.
A special thanks to Yves Schüler and colleagues for providing their financial cycle estimates.
For example, in the cast of the BIS estimates, we lose in total 366 observations, which is a little under 10% of our total number of observations.
The phase concordance index is a measure for co-movement between two cycles, say, f and b (see Harding and Pagan 2002). The corresponding expression for country i is given by \(\Phi ^{fb}_{i}=1/T\sum _t\left( \phi ^f_{it}\phi ^b_{it}+(1-\phi ^f_{it})(1-\phi ^b_{it})\right) \), where \(\phi ^{.}_{it}\) is 1 when the cycle is in an upturn and 0 when it is in a downturn as defined in the paper. If the phase concordance index equals 0 both cycles are perfectly counter-cyclical and when it equals 1 both cycles are perfectly pro-cyclical. If two cycles are independent we obtain the expected phase concordance, which can be obtained by replacing \(\phi ^{.}_{it}\) with fraction of time spent in an upturn for both cycles.
We have obtained higher concordance indices of around 0.6 between the Schüler cycle with a lead of two to eight quarters and the CPB cycle measures. A lead of two to eight quarters is actually short compared to the length of the financial cycle and there is still substantial overlap between the up- and down-phases of the Schüler and the CPB cycle.
When necessary we seasonally adjust the data using the X-12 algorithm.
Using other values for \(\lambda _{\textrm{HP}}\) yields qualitatively similar results.
We follow Hamilton (2018), who advises to use \(p=3\) and \(h=8\) for quarterly data.
If we include a short-term nominal interest rate (of the OECD’s Main Economic Indicators) instead of the real rate, all impulse responses are (roughly) unchanged aside from the interest rate. In this case the responses of the nominal rate display a resemblance of a standard Taylor rule for monetary policy with an increase following a rise in inflation, and a reduction after a drop in inflation.
We use the Fisher information matrix to calculate the standard error. However, because the log-likelihood is highly nonlinear, it is not certain that this results in a reliable standard error.
For the endogenous threshold 40% of the observations occur during an upturn of the financial cycle, while in the case of the exogenous threshold this percentage is 48%.
Austria, Germany, Italy, South Korea, The Netherlands, Sweden, and Switzerland show distinctly different patterns from the other countries and the overall main results. Denmark deviates to a lesser extent. For some of these countries, the data cover a shorter time period. However, it is also possible that an insufficient lag length, or a different underlying economic mechanism driving the interaction between the financial cycle and the investment multiplier might explain these results. We leave this for future research.
See footnote 16, replacing the real rate for a nominal rate does not change the results.
Germany, France, Japan, Norway, South Korea, Spain, The Netherlands, Switzerland and the USA show different patterns from the other countries and the overall main results (also see footnote 19).
To save space, the results are available upon request.
The results are also robust for the use of 2 lags, corresponding to 8 lags when using quarterly data. Since the information criteria only support the use of at most 6 lags when using quarterly data, there is no point in adding more lags to the model.
Changing the order of the endogenous variables leads indeed to identical results.
We find the same result with private GDP when using annual data.
Data are obtained from the World Bank over the period 1960–2016. The results are almost identical if we use the openness in 2016 instead of the entire sample.
Open economies: Belgium, Denmark, Ireland, the Netherlands, Norway, Sweden and Switzerland. The remaining eleven countries are closed economies.
Data obtained from Oxford Economics over the period 1980–2018.
Other orderings of the VAR model including debt lead to similar results.
Using the level of debt in 2017 instead of the entire sample, and using a cut-off of 100% such as Corsetti et al. (2012), yields similar results.
Highly indebted countries: Belgium, Canada, Denmark, France, Italy, Japan, the Netherlands and the USA. The remaining ten countries are all lowly indebted countries.
Belgium, Finland, France, Germany, Ireland, Italy, the Netherlands and Spain.
We have also estimated this truncated sample using an endogenous threshold. In this case the results are very similar to our baseline results.
Placing the financial cycle as endogenous variable in a different ordering does not change the results.
Their model also includes seasonal components to correct for any seasonality in the data. We omit this aspect of the model here for simplicity.
Note that the covariance matrices of both disturbance vectors \(\vec \kappa _{it}^{C}\) and \(\vec \kappa _{it}^{C*}\) are restricted to be equal. This restriction is standard, see Harvey (1991) for details.
Asymptotically, however, the two financial cycles will be qualitatively the same, only differing in their amplitude.
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Appendices
A State space model of CPB financial cycle
We denote the credit data for country i by \(y^1_{it}\) and the housing price index by \(y^2_{it}\) and define the data column vector \(\vec {y}_{it}=\left( y^1_{it}, y^2_{it}\right) '\). Similarly, we denote the vector of the financial cycles for credit and the housing price index of country i as \(\vec {f}_{it}\). We use the same notion for the vector of the business cycle components, \(\vec {b}_{it}\). The state space model also includes two local linear trend components, one for credit and one for the housing price index, giving rise to the vector \(\vec {\mu }_{it}\) to capture the growth dynamics in the data. Each trend component is itself affected by a random walk growth component, which for both series we also denote by a vector: \(\vec {\beta }_{it}\). Using this notation we can express the model in Luginbuhl et al. (2019) as followsFootnote 35:
where the financial cycle component is given by
and the business cycle component by
Note that the components \(\vec {b}_{it}^{*}\) and \(\vec {f}_{it}^{*}\) are only required for the construction of the cycle components \(\vec {b}_{it}\) and \(\vec {f}_{it}\). Also note that the model contains the vector disturbance terms \(\vec \varepsilon _{it}\), \(\vec \eta _{it}\), \(\vec \zeta _{it}\), as well as \(\vec \kappa _{it}^C \sim N\left( 0, \Omega _{\kappa ,i}^C \right) \) and \(\vec \kappa _{it}^{C*} \sim N\left( 0, \Omega _{\kappa ,i}^C \right) \) for \(C=b\) or f.Footnote 36
The state space model (Luginbuhl et al. 2019) use in the estimation of the financial cycle is closely related to work of Rünstler and Vlekke (2018) and Winter et al. (2017), which are also based on unobserved cycle components to model financial cycles. The main difference here being that the estimation method by Luginbuhl et al. (2019) imposes the restriction of there being only one shared stochastic process driving both financial cycle components in their bi-variate model. This restriction is necessary to help identify one financial cycle for each country. In fact the financial cycle estimate \(f_{it}\) is the cycle component for the credit data. Although the housing price index shares the same stochastic disturbance driving the cycle, the housing price index has its own starting value. As a result, the housing price index financial cycle can deviate from the credit cycle in the initial part of the sample period.Footnote 37
The bi-variate state space models distinguish between the business cycle and the financial cycle by assuming that the financial cycle has a longer period cycle than the business cycle. The financial cycles are full-information, Kalman smoothing estimates obtained using Bayesian Markov Chain Monte Carlo or MCMC methods.
B Data series
C Correlation and phase concordance between financial and business cycles
D IRFs of inflation and real interest rate, baseline specification
E Baseline results in alternative specifications
1.1 E.1 Figures for baseline results in alternative specifications
1.2 E.2 Alternative financial cycles
F Additional figures of the robustness checks
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Soederhuizen, B., Teulings, R. & Luginbuhl, R. Estimating the impact of the financial cycle on fiscal policy. Empir Econ 65, 2669–2709 (2023). https://doi.org/10.1007/s00181-023-02448-0
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DOI: https://doi.org/10.1007/s00181-023-02448-0