Abstract
Exchange rate uncertainty measured by its volatility is said to affect trade flows in either direction. A limited number of recent studies show that the response of trade flows to exchange rate volatility could be asymmetric, mostly due to change in traders’ expectations. In this paper, we test the symmetric and asymmetric effects of exchange rate volatility on Tunisia’s bilateral trade with each of its 16 partners. We find that Tunisia’s trade flows to each partner are affected asymmetrically in the short run but not in the long run. In almost half of the sample, the long-run effects were symmetric.
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Notes
Some recent studies since last review article are Hall et al. (2010) who considered the experience of emerging economies and Asteriou et al. (2016) who included MINT countries in their sample. Example of studies that use commodity level data include Bahmani-Oskooee and Hegerty (2009), (2011), Wong and Tang (2008), Bahmani-Oskooee et al. (2015), Bahmani-Oskooee and Durmaz (2016), Bahmani-Oskooee et al. (2017) and Belke et al. (2013, 2015).
The models and methods follow closely Bahmani-Oskooee and Aftab (2017).
If these income elasticities are negative, economic growth could be due to increased production of import substitute goods (Bahmani-Oskooee 1986).
Once the long-run estimates are normalized, we will have \(\frac{{\hat{\theta }_{1} }}{{ -\, \hat{\theta }_{0} }} = \hat{\alpha }_{1} ,\quad \frac{{\hat{\theta }_{2} }}{{ -\, \hat{\theta }_{0} }} = \hat{\alpha }_{2} \quad {\text{and }}\frac{{\hat{\theta }_{3} }}{{ - \,\hat{\theta }_{0} }} = \hat{\alpha }_{3}\) in (1) and \(\frac{{\hat{\rho }_{1} }}{{ - \,\hat{\rho }_{0} }} = \hat{\beta }_{1} ,\quad \frac{{\hat{\rho }_{2} }}{{ - \,\hat{\rho }_{0} }} = \hat{\beta }_{2} \quad {\text{and}}\quad \frac{{\hat{\rho }_{3} }}{{ -\, \hat{\rho }_{0} }} = \hat{\beta }_{3}\) in (2).
The third advantage of this method is that since short-run dynamic adjustment process is included in estimating long-run coefficients, the adjustment process allows feedback effects among variables to be accounted for and this reduces multicolinearity and endogeneity issues (Pesaran et al. 2001, p. 299).
Of course, this channel will be less effective if there exists a forward market for currencies that traders are holding.
The two partial sum variables are generated as \({\text{POS}}_{t} = \sum\nolimits_{j = 1}^{t} {\hbox{max} (\Delta \ln V_{j} } ,0),\quad {\text{NEG}}_{j} = \sum\nolimits_{j = 1}^{t} {\hbox{min} (\Delta { \ln }V_{j} ,0)} .\)
Note that Shin et al. (2014, p. 291) argue that the two partial sum variables in the nonlinear model should be treated as a single entry so that the critical values of the F test stay at high and conservative level when we move from the linear model to nonlinear model. For more on some other applications of these methods, see Halicioglu (2007, 2008), Kisswani and Nusair (2014), Gogas and Pragidis (2015), Durmaz (2015), Baghestani and Kherfi (2015), Al-Shayeb and Hatemi (2016), Lima et al. (2016), Aftab et al. (2017), Gregoriou (2017) and Bahmani-Oskooee et al. (2018, 2019).
There is now clear evidence that for small samples such as ours, the ARDL approach performs better than other approaches (Panopoulou and Pittis 2004).
Note that Banerjee et al. (1998) who introduced this test with Engle and Granger (1987) approach called this the t test for cointegration. Pesaran et al. (2001) extended it to ARDL model, and since variables could be combination of I(0) and I(1), they provide upper and lower bound critical values. However, the critical values are for large sample. For small samples such as ours, we rely upon Banerjee et al. (1998). For large samples, both sources report the same critical values.
Other diagnostic statistics are similar to those in Table 2 and need no repeat here.
Other diagnostics are similar to those in Table 6 and need no repeat here.
See also Zemami and Ben-Salha (2015).
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Appendix
Appendix
1.1 Data definitions and sources
Annual data over the period 1987–2016 are used to carry out the empirical analysis with each partner of Tunisia except that due to unavailability of some data, the study period was reduced to 1991–2016 for Germany. The data come from the following sources (Table 9):
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a.
Direction of Trade Statistics (DOT) of the IMF.
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b.
International Financial statistics (IFS) of the IMF.
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c.
OANDA Web site.
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d.
World Bank’s World Development Indicators.
1.1.1 Variables
\(X_{t}^{\text{TU}}\) = export volume of Tunisia to a partner. In the absence of export price index at bilateral level, following Bahmani-Oskooee and Hegerty (2009) we use and the aggregate export price index of Tunisia to deflate nominal exports. While nominal exports come from source a, export price index comes from source b.
\(M_{t}^{\text{TU}}\) = import volume of Tunisia from a partner. Again, in the absence of import price index at bilateral level, we relied upon Tunisia’s aggregate import price index to deflate nominal imports.
While nominal imports come from source a, import price index comes from source b.
YTU = Tunisia economic activity. We use real GDP as measure of economic activity. Data come from source d.
YP = Trading partner’s economic activity measured by real GDP. Data come from source d.
REX = The real bilateral exchange rate of the currency of a partner and Tunisian dinar. It is defined as (REX= PTUNEX/P) where NEX is the nominal exchange rate defined as number of units of a partner’s currency per Tunisian dinar (source c). PTU is the price level in Tunisia (measured by CPI) and P is the price level in a partner (also measured by CPI), source d. Thus, a decline in REX reflects a real depreciation of Tunisian dinar against a partner’s currency.
V = volatility measure of REX. Following Bahmani-Oskooee and Hegerty (2009), we use standard deviation of 12 monthly real bilateral exchange rates within each year as measure of volatility for that year. All monthly nominal exchange rates and monthly price level data come from source b.
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Bahmani-Oskooee, M., Nouira, R. On the impact of exchange rate volatility on Tunisia’s trade with 16 partners: an asymmetry analysis. Econ Change Restruct 53, 357–378 (2020). https://doi.org/10.1007/s10644-019-09250-y
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DOI: https://doi.org/10.1007/s10644-019-09250-y