Arora, Vivek, and Martin Cerisola, 2001, “How Does U.S. Monetary Policy Influence Sovereign Spreads in Emerging Markets?” IMF Staff Papers, International Monetary Fund, Vol. 48 (November), pp. 474–98.
Christofides, Charis, Christian Mulder, and Andrew Tiffin, 2003, “The Link Between Adherence to International Standards of Good Practice, Foreign Exchange Spreads, and Ratings,” IMF Working Paper 03/74 (Washington: International Monetary Fund).
Edwards, Sebastian, 1984, “LDC Foreign Borrowing and Default Risk: An Empirical Investigation,” American Economic Review, Vol. 74, pp. 726–34.
Eichengreen, Barry, and Ashoka Mody, 1998, “What Explains Changing Spreads on Emerging-Market Debt: Fundamentals or Market Sentiment?” NBER Working Paper No. 6408 (Cambridge: Massachusetts, National Bureau of Economic Research).
Ferrucci, Gianluigi, 2003, “Empirical Determinants of Emerging Market Economies’ Sovereign Bond Spreads,” Bank of England Working Paper No. 205 (London: Bank of England).
Glennerster, Rachel, and Yongseok Shin, 2003, “Is Transparency Good for You, and Can the IMF Help?” IMF Working Paper 03/132 (Washington: International Monetary Fund).
Institute for International Finance, IIF Action Plan Proposals and Dialogue with the Private Sector, 2002, Appendix D, “Does Subscription to the IMF’s Special Data Dissemination Standard Lower a Country’s Credit Spread?” (Washington: Institute for International Finance).
- Search Google Scholar
- Export Citation
)| false Institute for International Finance, IIF Action Plan Proposals and Dialogue with the Private Sector, 2002, Appendix D, “ Does Subscription to the IMF’s Special Data Dissemination Standard Lower a Country’s Credit Spread?” ( Washington: Institute for International Finance).
Kamin, Steven B., and Karsten von Kleist, 1999, “The Evolution and Determinants of Emerging Market Credit Spreads in the 1990s,” BIS Working Paper No. 68 (Basel: Bank for International Settlements).
MacKinnon, James G., 1996, “Numerical Distribution Functions for Unit Root and Cointegration Tests,” Journal of Applied Econometrics, Vol. 11, pp. 601–18.
Min, Hong G., 1998, “Determinants of Emerging Market Bond Spread: Do Economic Fundamentals Matter?” World Bank, Policy Research Paper No. 1899 (Washington: World Bank).
The author would like to thank J. R. Rosales, A. Pellechio, Carol S. Carson, Robert Flood, and colleagues in the IMF’s Statistics, International Capital Markets, and Policy and Development Review Departments for helpful comments and suggestions.
Complete information on the SDDS is available on the IMF’s Dissemination Standard Bulletin Board (DSBB), on the Internet at http://dsbb.imf.org/Applications/web/sddshome/.
The spread for a particular country is defined as its EMBI portfolio yield over a theoretical U.S. zero-coupon curve, where the sovereign yield is set to equate the total net present value of the sovereign risk cash flows to zero.
Early empirical investigations into the determinants of EME yield spreads found either no role for mature market interest rates (Min, 1998) or unexpected negative correlations (Eichengreen and Mody, 1998, and Kamin and von Kleist, 1999). More recently, Arora and Cerisola (2001) and Ferrucci (2003) report positive correlations between U.S. interest rates and EME spreads consistent with theoretical expectations.
The countries included in the panel were chosen to include those large emerging market countries subscribing to the SDDS with sufficient quarterly macroeconomic and bond issuance data available both before and after SDDS subscription to adequately inform the empirical work. On that basis, the basic panel includes data on Mexico, the Philippines, South Africa, and Turkey, covering the period 1990:1 to 2002:4, while data for Argentina, Brazil, and Colombia begin in 1994:1, 1992:4, and 1995:1, respectively, and cover the period up to and including 2002:4. Lack of data precluded the inclusion of many EMEs. In many cases, including for most of the transition countries, insufficient macroeconomic data (usually the absence of quarterly GDP estimates) prior to SDDS subscription precluded inclusion in the panel, despite several of these countries having benefited from access to private international capital markets. Certain other important EMEs, such as India and Singapore, did not issue any sovereign foreign currency-denominated bonds between 1990 and 2002.
And prior to the introduction of the euro in 1999, both deutsche mark- and ECU-denominated bonds.
Most foreign currency-bond issues are denominated in these three currencies. Additionally the BEL database does not calculate spreads for variable interest rate bond issues, regardless of the currency of denomination.
U.S. dollar-denominated bonds serve as the excluded category to preclude perfect multicollinearity.
The SDDS incorporated a formal transition period, beginning with the opening of subscription in early April 1996 and ending December 31, 1998. During this period Fund members could subscribe to the SDDS even if their dissemination practices were not fully in line with the SDDS at that time. This gave subscribers time to bring their data and dissemination practices into line with the standard according to a acknowledged transition plan.
Annual data for external debt (public and publicly guaranteed) stock-to-exports ratios, drawn from the World Bank’s GDF database, were converted to a quarterly frequency (same value for all quarters) then smoothed with the Hodrick-Prescott filter with standard quarterly parameters prior to testing the order of integration.
Tests with other pooled estimation techniques yielded broadly similar results.
Smoothed measures (four quarter moving averages) of real GDP growth rates and the two U.S. interest rates have been used in the regressions.
U.S. interest rates have been retained in the regressions, consistent with the fact that the U.S. dollar-denominated bonds represent the omitted category for other dummy variables. The correlation between the smoothed series USFED and USLONG for the period 1990:1 to 2002:4 is 0.64.
Ferrucci (2003) reports similar results, and suggests that this might be attributable to leveraged investors borrowing at short-term rates to lend at longer-term rates to EMEs. Tests with a single variable measuring the slope of the U.S. yield curve (10-year bond rate minus the 3-month Treasury yield) yielded a statistically significant negative coefficient estimate, but constraining the two coefficients to be equal but opposite in sign, proved overly restrictive and reduced the explanatory power of the regression.
These two coefficients test as statistically insignificantly from each other, and, at least in the spread specification, tests of their equality are readily accepted by the data.
Argentina: several reschedulings, including December 1989, September 1991, and July 1992, with the consolidation period of the latter ending in December 1994; Brazil: several reschedulings, with those of July 1988 and February 1992 falling within the sample period; Mexico: several reschedulings, including May 1989, with the consolidation period ending in May 1992; Philippines: several reschedulings, including May 1989 to July 1994, with the consolidation period of the latter ending in December 1995; Turkey: several reschedulings, from May 1979 to July 1980, but none during sample period. Colombia and South Africa have never rescheduled with Paris Club creditors.
All countries in the panel, save South Africa, have had financial arrangements with the IMF at some time over the course of the sample period.
A 90 percent confidence interval for this estimate ranges from -51 to -99 basis points.
Ideally, one would also wish to determine if increased capital market competition also played a role in reducing costs.
The GDDS also guides member countries in the dissemination of their economic and financial data to the public. Introduced in 1997, the GDDS is open to all IMF member counties, but not intended for those seeking access to international capital markets.