Abstract
Though not working toward an imminent transition to a monetary or currency union, the Central American Monetary Council (or CMCA, from Spanish Consejo Monetario Centroamericano) serves as an institution promoting economic and financial stability among five Central American countries (Costa Rica, El Salvador, Guatemala, Honduras and Nicaragua) and the Dominican Republic. Econometric studies conducted by researchers from CMCA have mostly focused on studying inflation levels of these countries, making use of econometric tools such as VECM and cointegration. We expand the study of inflation stability in the member countries of the CMCA by adopting a long memory and fractionally integrated approach and implementing cointegration methods that have not yet been used in the study of the Central American Monetary Council. Our results first show that all the series of prices are nonstationary, with orders of integration equal to or higher than 1 in all cases. Looking at long-run equilibrium relationships among the countries, we only found strong evidence of a cointegration relationship in the case of Honduras with El Salvador. All the other vis-a-vis relationships seem to diverge in the long run. Policy implications of the results obtained are also derived in the paper.
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Notes
Some authors argue that “mean reversion” is not a proper concept in the context of nonstationary series, i.e., if the fractional differencing parameter d belongs to the interval [0.5, 1). In this case, rather than “mean reversion” we can use the concept of “finite impulse responses” or “responses disappearing in the long run”.
The main advantage of the LM approach of Robinson (1994a) is that it remains valid even in nonstationary contexts (e.g., \(d=0.5\)). Moreover, it is the most efficient method in the Pitman sense against local departures from the null.
Classical cointegration as widely employed in the literature occurs then if \(d=1\) and \(b=1\).
Other more general definitions of fractional cointegration allow for different degrees of integration for each series (see, e.g., Marinucci and Robinson 2001). However, in a bivariate context, as is the case in the present paper, the two parent series must display the same degree of integration.
\(m=(n)^{0.5}\) is the bandwidth number usually employed in most empirical applications. Always, there is a trade-off between bias and variance: The asymptotic variance is decreasing with m, while the bias is growing with m.
This method requires covariance stationarity prior to the estimation. Thus, the values were estimated on the first differenced data, adding then 1 to the estimated values of d. We also performed the Abadir et al. (2007) approach that remains valid in the context of nonstationary series, and the results were completely in line with those reported with the tests of Robinson (1995a).
Non-fractional cointegration analysis was carried out by means of Engle–Granger (1987) cointegration tests. The results are displayed in Appendix 2, indicating that cointegration only takes places in three of the possible cases (Costa Rica with Nicaragua, Dominican Republic with Guatemala and Honduras with Guatemala).
We only display in Table 14 the values for \(m=15\) and 16. However, qualitatively the same results were obtained for other bandwidth numbers (m = 10, ..., 20) in terms of the unit root cases.
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Luis A. Gil-Alana acknowledges financial support from the Ministry of Education of Spain (ECO2014-55236). Comments from the Editor and two anonymous referees are gratefully acknowledged. This work was presented at the \(9\mathrm{th}\) Forum of Researchers from the CMCA at the Dominican Republic Central Bank on July 2015.
Appendices
Appendix 1: Unit root tests
ADF | Costa Rica | Dominican Republic | El Salvador | Guatemala | Honduras | Nicaragua |
---|---|---|---|---|---|---|
No regressors | 2.7257 | 2.6582 | 5.5378 | 7.0306 | 1.8294 | 5.4877 |
Intercept | 1.1572 | 0.43892 | \(-\)1.4980 | 1.3702 | 1.1991 | 2.7358 |
Intercept and Linear Trend | \(-\)1.9927 | \(-\)2.2145 | 2.1242 | \(-\)1.5862 | \(-\)1.1472 | \(-\)0.97942 |
PP | Costa Rica | Dominican Republic | El Salvador | Guatemala | Honduras | Nicaragua |
---|---|---|---|---|---|---|
No regressors | 11.9356 | 4.8872 | 5.9038 | 11.068 | 13.092 | 10.05 |
Intercept | 2.8245 | 1.1301 | \(-\)1.3870 | 1.6321 | 1.715 | 3.3581 |
Intercept and Linear Trend | \(-\)2.2658 | \(-\)2.0284 | \(-\)2.070 | \(-\)1.5785 | \(-\)1.960 | \(-\)0.7391 |
Appendix 2: Engle and Granger’s (1987) cointegration results
Dom. Rep. | Hond. | EL Salv. | Guatemala | Nicaragua | |
---|---|---|---|---|---|
COSTA RICA | \(-\)2.83 | \(-\)2.025 | \(-\)1.633 | \(-\)2.560 | \(-\) 3.362 |
DOM. REP. | XXXXX | \(-\)2.667 | \(-\)3.201 | \(-\) 3.933 | \(-\)2.478 |
HONDURAS | XXXXX | XXXXX | \(-\)2.159 | \(-\) 3.610 | \(-\)1.942 |
EL SALV. | XXXXX | XXXXX | XXXXX | \(-\)1.970 | \(-\)2.397 |
GUATEMALA | XXXXX | XXXXX | XXXXX | XXXXX | \(-\)3.227 |
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Carcel, H., Gil-Alana, L.A. Inflation analysis in the Central American Monetary Council. Empir Econ 54, 547–565 (2018). https://doi.org/10.1007/s00181-016-1223-0
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DOI: https://doi.org/10.1007/s00181-016-1223-0