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Erschienen in: Empirical Economics 2/2015

01.03.2015

DSGE priors for BVAR models

verfasst von: Thomai Filippeli, Konstantinos Theodoridis

Erschienen in: Empirical Economics | Ausgabe 2/2015

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Abstract

Similar to Ingram and Whiteman (J Monet Econ 34:497–510, 1994), De Jong et al. (in: Proceedings of the American Statistical Association Bayesian, 1993) and Negro and Schorfheide (Int Econ Rev 45:643–673, 2004) , this study proposes a methodology of constructing dynamic stochastic general equilibrium (DSGE) consistent prior distributions for Bayesian vector autoregressive (BVAR) models. The moments of the assumed Normal–Inverse–Wishart (no conjugate) prior distribution of the VAR parameter vector are derived using the results developed by Fernandez-Villaverde et al. (Am Econ Rev 97(1):21–26, 2007) , Christiano et al. (Assessing structural vars, 2006) and Ravenna (J Monet Econ 54(2):48–64, 2007) regarding structural VAR (SVAR) models and the normal prior density of the DSGE parameter vector. In line with the results from previous studies, BVAR models with theoretical priors seem to achieve forecasting performance that is comparable—if not better—to the one obtained using theory free ‘Minnesota’ priors (Doan, Econ Rev 3(1):1–100, 1984). Additionally, the marginal-likelihood of the time-series model with theory found priors—derived from the output of the Gibbs sampler—can be used to rank competing DSGE theories that aim to explain the same observed data (Geweke, Contemporary Bayesian econometrics and statistics, 2005). Finally, motivated by the work of Christiano et al. (Handbook of monetary economics, 2010a; Involuntary unemployment and the business cycle, 2010b) and Del Negro and Schorfheide (Int Econ Rev 45:643–673, 2004), we use the theoretical results developed by Chernozhukov and Hong (J Econom 115(2):293–346, 2003) and Theodoridis (An efficient minimum distance estimator for DSGE models, 2011) to derive the quasi-Bayesian posterior distribution of the DSGE parameter vector.

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Fußnoten
1
The term ‘full-information’ is loosely used here.
 
2
The only difference between the model described in this section and that developed by Smets and Wouters (2007) is that we have switched off the linear trend.
 
3
See Magnus and Neudecker (2002).
 
4
Similar to Ingram and Whiteman (1994) and De Jong et al. (1993), however, in contrast to Del Negro and Schorfheide (2004), our methodology does not offer a device of controlling the tightness of the prior such as the \(\lambda \) hyperparmeter in Del Negro and Schorfheide (2004). An anonymous referee suggested us a ‘way’ of extending our analysis to allow for such hyperparameter that we are going to investigate in a future work.
 
5
The exact steps of this Monte Carlo exercise are described in Appendix 1.
 
6
Instead of presenting 175 histograms—the total number of VAR parameters, \( 3\times 7^{2}+0.5\times 7\times (7+1)\)—that plot the prior distribution of the individual components \(\theta \left( \gamma _\mathrm{N}\right) \) against the elements of \(\theta \left( \gamma _\mathrm{GD}\right) \) we thought that it would be more constructive for the reader to present the impulse-response function as it best summarises all the VAR parameters. However, all the graphs are available from the authors upon request.
 
7
Canova and Sala (2009) call this type of estimators ‘full-information’ as it utilises all the available impulse-response information.
 
8
In Theodoridis (2011), the set of instruments employed for the estimation of the structural parameter vector correspond to the impulse-response function calculated using the OLS estimate of \(\theta \)\(\widehat{\mathrm{IRF}} \left( \hat{\theta }\right) \)—and \(\mathcal {W}\) is the inverse of the asymptotic variance–covariance matrix of \(\widehat{\mathrm{IRF}}\left( \hat{\theta } \right) \).
 
9
Similarly, Stock and Watson (2012) use the monetary policy shock from Smets and Wouters (2007) model to identify the policy shock in the VAR.
 
10
This data set is publicly available from the website of the American Economic Association.
 
11
This choice is consistent with the work of Smets and Wouters (2007).
 
12
Canova and Ferroni (2012) employ similar checks to assess the contribution of the prior moments to the posterior DSGE estimates.
 
13
Data dynamics are summarised by the impulse responses and they are influenced by the choice of the identification matrix.
 
14
The details about Minnesota priors can be found in Appendix 4.
 
15
The marginal-likelihood in this study has been approximated using Geweke’s-modified harmonic mean estimator (Geweke 1999). The calculation step can be found in either An and Schorfheide (2007) or Schorfheide (2000)
 
16
This is an interesting evidence that deserves further research that goes beyond the scope of this study.
 
17
The steps required to calculate these weights are described in Appendix 3.
 
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Metadaten
Titel
DSGE priors for BVAR models
verfasst von
Thomai Filippeli
Konstantinos Theodoridis
Publikationsdatum
01.03.2015
Verlag
Springer Berlin Heidelberg
Erschienen in
Empirical Economics / Ausgabe 2/2015
Print ISSN: 0377-7332
Elektronische ISSN: 1435-8921
DOI
https://doi.org/10.1007/s00181-013-0797-z

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