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We thank Andrew Berg, Patricia Brenner, Enzo Croce, Andrew Feltenstein, Daniel Hardy, Alain Ize, Meral Karasulu, Helene Poirson, Niam Sheridan, and Roberto Steiner for useful suggestions at different stages of this project, and to Jia Liu for her valuable research assistance. We also appreciate comments received by seminar attendees at the IMF Institute and MAE Department and at the 7th Annual Meeting of LACEA in Madrid, Spain.
The Asian crisis showed that this is true also for currencies accepted by a bloc of countries.
Developing economies may be particularly prone to this problem, as hedging opportunities are less available than in developed economies.
Dollarization of assets and liabilities in Mexico is subject to restrictions in the sample period.
We defined this variable according to the banking system aggregate we used for the dollarization and net foreign asset indicators. In most cases, since we used the broad banking system (Deposit Money Banks + Other Banking Institutions, or DMB + OBI), we defined LIABPS as the banking survey liabilities to the private sector. In those cases where we used DMB only, the relevant LIABPS was taken from the monetary survey.
In the case of Costa Rica, this is partly explained by unreported offshore operations by local banks.
These conclusions are preliminary, as some variables are being left out in our analysis, such as offshore operations.
In particular, one major difference arises in the case of Nicaragua, where throughout the sample period lending and deposit rates are essentially expressed in dollar terms, since “value maintenance” assured that interest rates were adjusted ex post by changes in the exchange rate.
These figures are definitely skewed upward as a result of the hyperinflation years of the early 1990s, where nominal interest rates and spreads became extremely high and volatile. Taking only the 1995-2001 period, the average domestic currency relative spread falls to 20 percent and its standard deviation falls to 2 percentage points, while the foreign currency counterpart falls only slightly, to just under 10 percent.
The IFS exchange rate value for Argentina registers very small changes around 1.000 during 1991 to 1995, therefore INTERV is not strictly equal to unity in this period although it is clear that a hard peg was in place.
Specific indices used are reported in the following subsection.
In earlier regressions for quarterly data we also used a second indicator of banking system development, private sector credit as a percentage of banking system liabilities to the private sector. Given the variability of the denominator, this variable did not perform as well as FIN in predicting dollarization of liabilities.
It was not possible to find adequate measures of other structural variables, such as indicators of exchange controls or prudential regulation.
We also used a second measure, which averaged INTERV over the previous 36 months, with similar results.
This was obtained from the World Bank database on deposit insurance systems around the world.
We also undertook estimations using an alternative indicator, COVGE 1, a dummy variable taking a value of one when a country had a high level of deposit insurance coverage (greater than three times the GDP, roughly the world average), and zero otherwise. The results were similar to those obtained with COVGE.
For inflation = INF and the bilateral real effective exchange rate with the U.S. = REERUS, DOLINDEX is defined as:
[VAR(INF)+COV(INF, REERUS)]/[VAR(INF)+VAR(REERUS)+2 COV(INF, REERUS)].
There may be double counting for lending by branches operating domestically.
This result holds with either measure of openness: exports/GDP (reported) or exports plus imports/GDP.
Because of this puzzling result, equations 3, 6, 9, 12, 15, 18, 21, 24, 27, and 30 report results excluding deposit insurance coverage as an explanatory variable.
We excluded Haiti and Honduras altogether, because they show relatively few annual observations.
One alternative explanation may be that deposit insurance coverage summarizes past experiences of bank bailouts. If this were the case, the negative sign would show that in countries where sizable bailouts have already taken place, moral hazard incentives diminish along time. Another explanation, although affecting only one observation in the sample, could be that the one time variation captured by the regression—the adoption of deposit insurance in Jamaica in 1998—may have actually corresponded to a reduction in moral hazard, to the extent that a limited explicit scheme replaced a previous unlimited implicit scheme. This type of phenomenon was observed in one study of the adoption of deposit insurance in the EU (Gropp and Vesala, 2000).