Abstract
There is no a priori reason to suppose that price-setting behaviour is homogeneous across sectors and countries. Aggregate data are, however, commonly used to estimate the New Keynesian Phillips curve (NKPC), which may very well yield erroneous results if price-setting behaviour is heterogeneous. In this paper, we therefore estimate the hybrid NKPC for the Euro Area using a novel sectoral data set containing quarterly observations from 1999Q1 to 2012Q1. We show that a positive relationship between inflation and real marginal cost cannot be established empirically for a majority of countries and sectors. We also perform a meta-analysis by combining the results of individual significance tests in order to assess the validity of the NKPC in each country across all sectors and in each sector across all countries. We find no empirical evidence for the NKPC in the Euro Area when this meta-analysis is used. Our results therefore raise doubts about the appropriateness of the NKPC for the analysis of inflation dynamics and monetary policy in the Euro Area.
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Notes
The hybrid version of NKPC is, however, not completely based on micro-foundations as it does not evolve from optimization problem. The inclusion of lagged inflation into the NKPC is therefore justified by using an ad hoc rule for price-setting behaviour as suggested by Galí and Gertler (1999). An alternative way to rationalize the presence of lagged inflation is the price indexation rule proposed by Christiano et al. (2005).
The condition \(\lambda _{ij}^f +\lambda _{ij}^b <1\) is required to ensure that \(\left| {\rho _{ij,1}} \right| <1\) and \(\left| {\rho _{ij,2}} \right| >1\). In this case, a unique rational expectation solution exists and inflation is stationary. If \(\lambda _{ij}^f +\lambda _{ij}^b =1\) and \(\lambda _{ij}^f <0.5\), then \(\rho _{ij,1} =1\) implies that inflation is \(I\left( 1 \right) \) since \(s_{ij,t} \sim I\left( 0 \right) \) by construction. However, if \(\lambda _{ij}^f +\lambda _{ij}^b >1\), then a unique rational expectations solution does not exist (Nymoen et al. 2012; Dees et al. 2009).
Such as arts, entertainment, recreation, activities of households and extraterritorial organizations.
We refer to Petrella and Santoro (2012) for a detailed theoretical explanation on why the sector-specific labour income share is considered as a suitable proxy for the real marginal cost.
The smoothing parameter for the Hodrick–Prescott filter is set to 1,600, which is typically used for quarterly data.
The maximum number of factors is set to eight times the smallest integer that is larger than \(\left( {T/100} \right) ^{0.25}\).
We have also estimated the sectoral hybrid NKPC using data that has not been corrected for outliers. We found that the results are rather sensitive to the presence of irregularities in the data. However, we found that even fewer cases yield statistically significant parameter estimates when outliers are not removed from the data, which is a conservative result when related to our findings.
The data are seasonally adjusted using a seasonal dummy variables regression.
The BSQT method does not lead to a single rejection for unemployment and government expenditure as a ratio to GDP. These variables, however, are not excluded from the data set since according to economic theory they are considered to be mean reverting processes; therefore, the failure to reject the null hypothesis of a unit root can be attributed to inefficiency due to small sample size.
The estimation results of the structural parameters for sectoral hybrid NKPC, namely the subjective discount factor \(\beta _{ij}\), the fraction of backward-looking price setters \(\omega _{ij}\) and the rigidity parameter \(\delta _{ij}\) are not reported in this paper. These results are omitted due to convergence problems that have resulted in very imprecise structural parameter estimates, but they are available upon request. We have also performed the nonlinear version of Anderson–Rubin test in order to test the null hypothesis that the structural parameter estimates are admissible to the data (we refer to Stock et al. 2002; Ahrens and Sacht 2014 for a general discussion on this test). The results of Anderson–Rubin test show that the null hypothesis of weak identification is rejected in most of the cases, implying that the structural estimation results should be interpreted with caution. The Anderson–Rubin test results are available upon request.
As a robustness exercise, we have estimated the hybrid NKPC using aggregated data for each country and the Euro Area as a whole. The results, however, provide no statistical support for the hybrid NKPC when aggregated data are used, implying that inferior performance of the aggregate hybrid NKPC might be attributed to implicit aggregation bias. These results are available upon request.
We have also estimated the pure forward-looking sectoral NKPCs, which are not reliant on the assumption of a rule-of-thumb price-setting behaviour as in Galí and Gertler (1999). In order to avoid the misspecification bias, we only consider the cases for which the backwardness parameters \(\lambda _{ij}^b \) are statistically insignificant. These results show that the forward-looking sectoral NKPC is not supported by the data in all the cases. The results are available upon request.
In each individual test, a probability to reject erroneously the null hypothesis corresponds to significance level \(\alpha \). As an example, a probability for at least one false rejection in \(n\) individual independent tests is \(1-\left( {1-\alpha } \right) ^{n}\). In this case, at the 5 % significance level, the FWER is \(1-\left( {1-0.05} \right) ^{13}=0.487\) and \(1-\left( {1-0.05} \right) ^{5}=0.226\) when joint significance is tested across all sectors and countries, respectively, which are obviously larger than the desired overall significance level of 0.05.
For an extensive discussion about wavelet analysis, we refer to Gençay et al. (2002).
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Acknowledgments
The author would like to thank Fredrik N.G. Andersson, David Edgerton, Peter Karpestam, Yushu Li, Joakim Westerlund and two anonymous referees for constructive comments and suggestions. A previous version of the paper was presented at a seminar at Lund University and at the seventh Nordic Econometric Meeting in Bergen, Norway. Financial support from the Jan Wallander and Tom Hedelius Foundation under research Grant Numbers P2009-0189:1 and P2014-0112:1 is gratefully acknowledged.
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Appendices
Appendix 1
See Tables 8, 9, 10, 11, 12 and 13.
Appendix 2: A wavelet-based procedure for outliers correction
The data are corrected for additive outliers employing a procedure based on wavelet analysis. The principal idea is to decompose the time series into different frequency components, which is in essence a band pass filtering.Footnote 16
We transform the data using a maximal overlap discrete wavelet transform (MODWT). The first-level wavelet coefficients correspond to the vector of detail coefficients \({\mathbf {D}}_1 =\left( {d_{1,1}, d_{1,2}, \ldots d_{1,N}} \right) \) and the approximation coefficients \({\mathbf {A}}_1 =\left( {a_{1,1}, a_{1,2}, \ldots , a_{1,N}}\right) \), where \(N\) is the number of wavelet coefficients. We use the Haar wavelet filter as there is no need to employ a smoother wavelet filter for outlier detection (Granè and Veiga 2010).
The first-level detail coefficients \({\mathbf {D}}_1\) are very sensitive to any kind of abrupt changes in data, such as presence of outliers. Therefore, it is sufficient to screen only the finest-scale detail coefficients (Bilen and Huzurbazar 2002; Granè and Veiga 2010). Following Bilen and Huzurbazar (2002), the detail coefficients in \({\mathbf {D}}_1\) are compared with a threshold limit given by
where \(\hat{\sigma }_1\) is the mean of the absolute deviations from the median based on the finest-scale detail coefficients
where \(M_1\) is the median of \({\mathbf {D}}_1\). Outlier correction is carried out by setting to zero those detail coefficients which exceed the threshold limit \(\tau _1\). Formally, a hard-thresholding procedure is performed and the modified finest-scale detail coefficients \(\tilde{{\mathbf {D}}}_1 =\left( {\tilde{d}_{1,1},\tilde{d}_{1,2},\ldots \tilde{d}_{1,N}}\right) \) are obtained as \(\tilde{d}_{1,n} =0\) if \(\left| {d_{1,n}} \right| >\tau _1 \) and \(\tilde{d}_{1,n} =d_{1,n} \) otherwise for \(n=1,\ldots ,N\). Finally, the time series are reconstructed by applying inverse maximal overlap discrete wavelet transform (IMODWT) using the modified detail coefficients \(\tilde{{\mathbf {D}}}_1\) and the original approximation coefficients \({\mathbf {A}}_1 \).
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Norkute, M. Can the sectoral New Keynesian Phillips curve explain inflation dynamics in the Euro Area?. Empir Econ 49, 1191–1216 (2015). https://doi.org/10.1007/s00181-014-0909-4
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DOI: https://doi.org/10.1007/s00181-014-0909-4