Abstract
The literature highlights that inflation targeting (IT) economies enjoy a sizeable share of global trade, suggesting IT is a comparative advantage exclusive to such countries. Using a set of developed and developing countries—coupled with synthetic control methods—we estimate the causal effect of IT on the volume of international trade. We compare the post-IT path of each treated country to its synthetic counterpart. We find that IT adoption had sizeable positive (negative) long-run effects on Mexico’s (Brazil’s) trade and short-run effects, as large as 11 percentage points, on Uganda’s trade. For treated countries in Asia, Thailand recorded increases in trade, whereas the Korea Republic experienced nonlinear IT-induced trade effects. Norway registered negative IT effects, and Sweden, Canada, New Zealand and the UK experienced nonlinear effects in the post-IT period. IT induced negative and positive import and export volume and price changes in additional targeters. These heterogeneous IT-induced changes suggest that IT is not a source of comparative advantage in trade for all targeters and all post-adoption years.
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Notes
Greenspan (1994) defines price stability as the growth rate of inflation being sufficiently low such that households and firms do not consider it when making decisions.
See Svensson (2010) for a discussion on the history, macroeconomic effects, theory, practice, and future of inflation targeting.
Broadly speaking, we use Anderson and van Wincoop’s (2004) definition of trade costs, which includes all costs incurred in moving a good from producer to final user.
In the 2015 joint OECD-WTO Aid for Trade monitoring survey, 87.0% of the 62 developing and least developed country respondents indicate that trade costs are very important for their export competitiveness, whereas 91.9% are of the view that trade costs are important or very important for access to imports.
Allowing the control group to include suitable countries across all regions increases the size and power of the test Billmeier and Nannicini (2013).
In terms of bias and variance, Ferman and Pinto (2019) show that the synthetic control estimator fares better over its DID counterpart in settings where treatment assignment is correlated with stationary common factors, non-stationary trends are present in the data, an imperfect pre-treatment fit exists, and the number of pre-treatment periods is not large.
Additionally, the SCM limits study design bias in that researchers can construct a model unaware of how their design would influence the outcome of their study Abadie et al. (2010).
Nevertheless, if the IT framework improves a country’s competitiveness in global trade, given that imports and exports are elastic to price changes, there may be a decrease in imports and an increase in exports; total trade may, hence, change based on the greater magnitude of these opposing effects.
For example, Finland, Spain and the Slovak Republic adopted IT but eventually abandoned the regime.
We thank two anonymous referees for highlighting these points.
Our work also sits with the applied strand of SCM that investigates treatment effects in different macroeconomic contexts such as the evaluation of the impact of terrorism (Abadie and Gardeazabal 2003), economic and trade liberalization (Billmeier and Nannicini 2013), natural disasters (Cavallo et al. 2013). Ferman et al. (2018) document uses of the SCM in comparative case studies of civil wars and political risk, natural resources, international finance, education and research policy, health policy, political reforms, labour, taxation, crime, social connections, and local development.
Abadie and Gardeazabal (2003) indicate that the restrictions on the weights in \(\tau \) prevents extrapolation bias thus restricting extrapolation within the support of our trade covariates for the control group. The vector of covariates for the treated unit will only be perfectly fitted if it lies in the convex hull of the trade covariates for the control group.
Moreover, given the prevailing pattern in the adoption of IT regime by some countries, prima facie evidence to strongly suggest that the policy of IT adoption is contagious across countries is lacking.
Note that an abundance of missing observations precludes us from starting prior to 1984.
The length of such period is strictly left to our discretion as no reading thus far has indicated a prescribed length.
Although the GDP deflator and the CPI are measures of inflation, note that an IT regime is predicated on the CPI measure of inflation and thus cannot be included in the model specification. Also, our SCM results with and without the GDP deflator are quantitatively and qualitatively indistinguishable.
In the sensitivity analyses, we perturb the core with additional covariates such as regional and oil exporter indicators, population, arable land, general government consumption and official exchange rate, which were also retrieved from The World Bank (2019). We also control for institutional quality, which is proxied by an average of variables from the International Country Risk Guide Database.
In practice, two main data conditions must be satisfied for the SCM to work. Firstly, no observation must be missing for the outcome variable. Secondly, none of the covariates should have missing observations for the entire pre-treatment span of the treated unit under observation. Due to these two conditions, many countries were excluded from the sample. Following Abadie et al. (2010), we also exclude from the donor pool all countries that adopted IT but eventually abandoned the regime; these countries include Finland, Spain and the Slovak Republic. This was done because the counterfactual outcome is meant to reproduce what the outcome would be in the absence of IT.
It is quite common in SC applications to have the number of control units exceeding the number of pre-treatment periods. Furthermore, Ferman (2019) shows that the synthetic control estimator can be asymptotically unbiased in settings where the number of control units exceeds the number of pre-treatment periods.
All excluded information can be furnished upon request
All control units contributed to the development of Canada, New Zealand and the UK synthetic trade.
For some targeters, there seems to be a divergence between actual and synthetic trade in the post-IT short run; however, these trajectories signal some level of convergence in the long run.
For example, for Korea Republic, the IT effect ranges from 26.18 to 51.34 percentage points from 2008 to 2017 but was \(-\) 14.60 (\(-\) 14.82) percentage points in 2002 (2003).
Having implemented the policy in 2000 and 2001, respectively, both countries’ pre-IT counterfactual trade fit the observed relatively well, with South Africa boasting an RMSPE and \({\bar{R}}^2\) of 0.81 and 1, respectively, relative to Uganda’s values of 1.98 and 0.96. Synthetic South Africa is principally made up of Argentina (42.2%), Mauritania (10.3%) and Togo (12.4%). Synthetic Uganda’s primary donors are Bangladesh (65.1%) and Sudan (13.6%).
The pattern of long-run convergence between actual and counterfactual trade is also evident among these targeters.
For Colombia, the significant IT effect on trade ranges from \(-\) 26.56 to \(-\) 16.51 percentage points from 2009 to 2014; for Brazil, the IT effect ranges from \(-\) 20.10 to \(-\) 12.69 percentage points from 2005 to 2014; and for Mexico, the IT effect ranges from 17.93 to 39.06 percentage points from 2013 to 2017. Many of these effects are beyond the 5-year horizon, suggesting that medium- to long-run IT effects on trade may be more sensitive to pre-treatment fit.
Stockman (1985) indicates that fully anticipated inflation influences the cost of holding money, which may have real effects on the exchange rate and international trade flow.
Anecdotes suggest that the length of the transitional period differs across countries and depends primarily on the time taken to integrate the key elements. Albania, for example, adopted an IT lite regime in 2004 and later shifted to a full-fledged version in 2007, seeing a transitional period of three years. Additionally, the works of Ltaifa et al. (2012) and Pruski (2004) highlight that Brazil, Chile, Czech Republic, Ghana, Hungary, Israel, South Africa and Turkey experienced a transitional period of six months, ten, one, five, eight, two and four years, respectively. Thus, we choose the average transitional period to be four years.
In reconciling these results with those of the synth wrapper in Table 8, we note that the synth wrapper results are adjusted for pre-treatment fit. However, the p-values we report for the null hypothesis of no IT effect on trade do not enjoy such adjustment.
Thailand’s sizeable estimated effects, coupled with its in-time p-value and its trade dynamics with China during the post-IT period, warrant further investigation. Thus, we re-estimate Thailand’s SCM results by setting the timing to 4 years after IT implementation. We find that Thailand experiences material, positive and statistically significant IT effects on its trade in its post-IT period. However, these effects are smaller in magnitude than their counterparts implied by the in-space placebo. We thank a reviewer for pointing this out.
Galiani and Quistorff (2017) divide all SCM effects by the corresponding pre-treatment match quality, as measured by RMSPE, to get “pseudo t-statistic” measures and “pseudo p-values”.
IT countries omitted from this table register insignificant effects in the synth-wrapper.
These estimates are from the synth wrapper that reports significance for more than 50% of the post-IT period-specific SCM estimates.
We do not simultaneously control for all covariates because this would render our effective sample size useless.
With the introduction of the logarithm of arable land in the augmented model, Sudan was ejected from the control group due to missing observations in the pre-treatment period for all targeters.
In some instances, its inclusion marginally increases the RMSPE in the pre-treatment period for targeters in which Sudan had previously contributed to their synthetic outcomes; these countries include Albania, Brazil, India, Korea Republic, Mexico, South Africa, Sweden, Uganda, Uruguay.
See Auboin and Ruta (2013) for a recent review.
The addition of the official exchange rate to the augmented model resulted in Austria, Puerto Rico, France, Honduras, Malta, Mongolia and the Netherlands exiting the control group.
This sample excludes China.
With the addition of government consumption, 8 countries exited the control group.
Also, there were marginally increases in the pre-IT RMSPE for Chile, Colombia, Brazil, Ghana, Guatemala, India, Indonesia, Korea Republic, Mexico, Norway, South Africa, Sweden, Thailand, Turkey, Uganda, Uruguay.
The literature has examined the effect of various aspects of institutional quality on trade. See, for example, Anderson and van Wincoop (2003); Melitz (2003); Dunlevy (2006); Berkowitz et al. (2006); Aviat and Coeurdacier (2007); Nunn (2007); Levchenko (2007); Head et al. (2010); Amiti and Weinstein (2011); Levchenko (2012); Briggs (2013); Ferguson and Formai (2013); Sequeira and Djankov (2014); Araujo et al. (2016).
The addition of the index to the augmented trade model resulted in 16 countries exiting the door pool.
The latter manifests when the characteristics of the experimental unit are incorrectly parallelized by merging idiosyncratic differences in sizeable group of untreated units.
We thank a reviewer for suggesting this idea.
We thank the reviewers for highlighting this point.
We use the synth wrapper to generate period-specific SCM estimates. The words in parentheses indicate the direction of the estimated IT effect.
These import and export price data are from the Penn World Tables.
For only a few post-IT years, Norway, Dominica Republic and Uganda experienced changes in their export prices, whereas Paraguay, Uruguay, Turkey and South Africa realized changes in their import prices.
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We are extremely grateful to the Editor, an Associate Editor and two anonymous referees for valuable comments that helped to significantly improve our paper’s quality. Also, we thank seminar participants in the Department of Economics at The University of the West Indies at Mona for their comments. All errors are our own The views expressed are those of the authors and do not necessarily reflect those of their institutions.
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McCloud, N., Taylor, A. Does inflation targeting matter for international trade? A synthetic control analysis. Empir Econ 63, 2427–2478 (2022). https://doi.org/10.1007/s00181-022-02221-9
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DOI: https://doi.org/10.1007/s00181-022-02221-9
Keywords
- International trade
- Import prices
- Export prices
- Inflation targeting
- Synthetic control methods
- Treatment effect