Abstract
The present paper analyzes how shocks to the copper price affect the economy of Chile, the world’s largest producer of this metal. Chile is an interesting case study, it being an emerging commodity producing economy that has a fiscal rule, which explicitly takes into account the expected future copper price, and has adopted an inflation targeting monetary policy with a fully floating exchange rate. The empirical analysis consists in estimating structural vector autoregressive models where the shocks are identified by sign restrictions in order to make the distinctions of those caused by increasing world demand, decreasing copper supply, and specific copper demand, e.g., speculation in future price increases. Positive copper price shocks, independently of the source, result in an appreciated currency, whereas the effects on the other macroeconomic variables do depend on the source of the shock. While demand shocks affect the Chilean gross domestic product positively, this is not the case when the copper price increases because of a supply-side event or a specific copper demand. Particularly, the activity in the mining sector is affected, while the non-mining sector expands only when the shock is caused by increasing world demand. One explanation could be the stabilizing effect of the fiscal rule. The impacts on inflation and the interest rate are correlated because of the inflation targeting monetary policy.
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Notes
For oil price shocks this has been documented by e.g. Kilian (2009).
For an early example, see Hotelling (1931).
Van der Ploeg (2011) provides a discussion of why the presence of natural resource is a benefit for some countries, while this is not the case for others.
Chen and Rogoff (2003) argue that the commodity’s price has a strong influence on the real exchange rate in Australia, Canada and New Zealand. Cashin et al. (2004) find a long-run relation between the real exchange rate and real commodity prices in one-third of the countries in their sample of 58 commodity-dependent countries. Chile, however, is not one of them. Chen et al. (2010) provide evidence that real exchange rates in commodity exporting countries are higher in periods where commodity prices are high.
Chapter 4: Commodity Price Swings and Commodity Exporters.
These numbers exclude eastern bloc countries.
These figures are only for primary supply. An important part of the supply is secondary, i.e. recycling of copper.
The main purpose of copper (almost 70% of total use in 2012) is in the manufacturing of electric and electronic products and the construction of buildings. In the present analysis, the copper price used is that of refined copper and world demand is measured by the gross domestic product (GDP), which includes the construction sector, rather than industrial production, which is the indicator frequently applied in oil price studies.
Numbers cited are from Meller (2013).
With observations from 1948 to 1968, Fisher et al. (1972) argue that the Chilean mining sector is hardly sensitive to price changes. Vial (2004) employs 30 years of data and finds for Chile a small and negative short-run price elasticity for the copper production, while it is positive in the long run.
Even though the monetary policy in an inflation targeting economy should, per se, not react to exchange rate movements, it does react to the impact these changes have on inflation and, particularly, second-round effects. A recent study on the exchange rate pass-through in Chile is that of Justel and Sansone (2015).
This methodology was introduced in the late 1990s by Faust (1998), Canova and De Nicoló (2002, 2003), and Uhlig (2005), and by now several empirical applications with this approach have been published in academic journals. Fry and Pagan (2011) present a critical review of the sign restriction method.
Identification should be improved if the all restrictions are satisfied at the same time (Paustian 2007).
It should be noted that the fiscal rule in Chile has been operative since 2000, though with some changes across time, and the exchange rate has been freely floating since 1999, where inflation targeting was fully implemented. Hence, the sample analyzed contains only few observations before the introductions of these regimes, which should not affect the results.
The GDP level series extracted from the June 2013 CD-issue of IFS contains observations till 2012Q1, but is discontinued. The growth rate series utilized is from the December 2016 issue. Before 2008Q1 the revisions of the growth rates are minor (less than 0.2 basis points), hence the series is updated with changes from that quarter onward.
Own translations: “Total employment”, “Employment by economic sector. Exploration of mining and quarrying” and “Table 3—spliced series”.
Furthermore, if the real interest rate and the inflation rate are stationary, the nominal interest rate must be stationary as well. A similar specification is applied by Peersman and Van Robays (2009) for studying the impact of oil price shocks.
To defend the exchange rate, the Central Bank of Chile raised the policy rate from 8.5 to 14% in September 1998.
The hypothesis of excluding the deterministic trend could not be rejected with a p value of 0.14. On the other hand, the hypotheses of excluding the seasonal dummies and the blip dummy were both rejected with p values of 0.
Fornero and Kirchner (2014) examine a DSGE model for Chile where agents learn about the persistence of commodity price shocks. They find that investment increases in a gradual way driven by investment in the commodity sector. In the same line, Fornero et al. (2014) consider six major commodity exporting countries and argue that in most of them the expansionary effects of positive commodity price shocks are driven by investment in the commodity sector and there is spillover to the other sectors.
The nominal copper price is deflated with the US producer price index extracted from the IMF’s IFS database. The correlation between nominal and real prices is 0.99 for the period considered.
The MTs presented do not change if the variability measure is calculated as the average variance instead of the maximum.
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Acknowledgements
I appreciate the insightful comments and suggestions from two anonymous reviewers of this journal and from Joaquín Vial as well as participants at the 17th World Congress of the International Economic Association, 20th International Conference on Computing in Economics and Finance, 2014 Annual Meeting of the Chilean Economic Society, XIX Annual Meeting of the Central Bank Researchers Network and seminars organized by Central Bank of Chile and Universidad de Santiago de Chile. I am grateful to Camila Figueroa for excellent research assistance. The usual disclaimers apply and the views and conclusions expressed in this paper do not necessarily represent those of the Central Bank of Chile or its board members.
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Appendices
Appendix A: MT impulse response function
The MT impulse response functions of Fry and Pagan (2011) corresponds to the impulse response function of the individual structural model, out of the 1000 estimated, which is element-wise closest to the median. In the present context, the MT impulse response functions are defined by optimizing with respect to the responses of the local variables from impulses to the global ones, i.e., responses of four variables from three shocks, for 20 quarters. Let \( \hat{\varTheta }_{q,n}^{{\left( {r,s} \right)}} \) be the response of variable r = 1, …, 4 for the quarter q = 1, …, 20 to the shock s = 1, 2, 3 of the successful draw n = 1, …, 1000, and let \( \hat{\varTheta }_{{q,{\text{med}}}}^{{\left( {r,s} \right)}} \) denote the median response of variable r for quarter q from a shock s. The MT impulse response is calculated as:
where the measure of variability of the set of successful impulse responses for variable r and shock s is calculated as \( \hat{V}_{r,s} = \mathop {\hbox{max} }\nolimits_{{q \in \{ 1, \ldots ,20]}} \frac{1}{1000}\mathop \sum \nolimits_{n = 1}^{1000} \left( {\hat{\varTheta }_{q,n}^{{\left( {r,s} \right)}} - \left( {\frac{1}{1000}\mathop \sum \nolimits_{n = 1}^{1000} \hat{\varTheta }_{q,n}^{{\left( {r,s} \right)}} } \right)} \right)^{2} \).Footnote 35
Appendix B: Preliminary tests
Appendix C: Robustness analyses
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Pedersen, M. The impact of commodity price shocks in a copper-rich economy: the case of Chile. Empir Econ 57, 1291–1318 (2019). https://doi.org/10.1007/s00181-018-1485-9
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DOI: https://doi.org/10.1007/s00181-018-1485-9