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The Ties that Bind: Measuring International Bond Spillovers Using Inflation-Indexed Bond Yields

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Abstract

This paper explores international bond spillovers using daily and weekly data on yields on inflation-indexed bonds and associated inflation expectations for the United States, Australia, Canada, France, Sweden, Japan, and the United Kingdom. The analysis starts in 2002, by which point U.S. inflation-indexed markets had matured. Real bond yields are found to be closely linked across countries, with developments in U.S. markets determining around half of real foreign yields and no evidence of spillovers back to the United States. Spillovers in inflation expectations are smaller and the direction of causation is less clear.

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Notes

  1. See Sack and Elsasser (2004) for an overview of the U.S. inflation-indexed market.

  2. Jorion (1996) and Breedon, Henry, and Williams (1999), who examine longer maturities, find no evidence of real interest parity across the major industrial countries. Chinn and Frankel (1995) find little evidence of real interest parity for shorter maturities. Gagnon and Unferth (1995); Goodwin and Grennes (1994); and Awad and Goodwin (1998) do find some support for real interest parity for short-term interest rates. See Ferreira and León-Ledesma (2007) for recent evidence on real interest parity for both developed and emerging economies, and for additional references.

  3. The yield on an inflation-indexed bond would also include a premium to compensate investors for its lower liquidity relative to a conventional bond. See Sack (2002) and Shen (2006) for estimates of the size of this liquidity premium.

  4. Sweden issued its first inflation-indexed bond in March 1994.

  5. Germany first issued an inflation-indexed bond in 2006. The yields are also highly correlated with French yields.

  6. The only significant deviation is in the case of Canada, for which the first inflation-linked bond matures in 2021 and is used throughout the sample.

  7. Analysis using earlier start dates—1998 and mid-2000—find very similar results except that, as expected, there is more evidence of inefficiencies in the U.S. inflation-indexed market (and in derived inflation expectations).

  8. Correlations since mid-2000 do not show this anomaly.

  9. This is true irrespective of the links across markets. For example, bond yields could be cointegrated if they react to information in a similar manner. Even so, past bond yields should not matter for current movements.

  10. Note that there is a small expected change in the return due to the shift in the duration of the bond from t to t + 1, but with a 10-year bond, a change of one day makes no noticeable difference.

  11. U.S. bonds are unique in that they are traded more or less continually 24 hours a day, and (in contrast to the other countries in the sample) the fix in our data set varies slightly depending on the day, varying from 3 to 7.30 pm EST.

  12. Note that, since interest rates are a random walk, the presence of cointegration would not influence these results. In any case, standard tests show little evidence of cointegration.

  13. We also ran rUS, p*, r*, pUS for all countries but the results were quite similar to rUS, r*, p*, pUS.

  14. These results are consistent with Chinn and Frankel (2005) and Cumby and Mishkin (1986).

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Authors
  1. Tamim Bayoumi
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  2. Andrew Swiston
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Additional information

*Tamim Bayoumi is a senior advisor with the IMF Strategy, Policy, and Review Department, and Andrew Swiston is an economist with the IMF Western Hemisphere Department. The authors gratefully acknowledge extremely helpful comments on this paper from two seminars at the IMF.

APPENDIX

APPENDIX

Figures A1, A2, A3, A4, A5, A6, A7, A8, A9, A10, A11 and A12 show impulse-response functions up to a horizon of 50 days. Interest rates and inflation expectations are expressed in percentage points. The figures show the response, in percentage points, of the first variable listed to a one standard deviation shock in the second variable listed. The ordering of the Cholesky decomposition used to identify shocks to each variable is RUS, PUS, R*, P* in Figures A1, A2, A3, A4, A5 and A6; RUS, R*, P*, PUS in A8, A9, A10, A11 and A12; and R*, P*, RUS, PUS in A7, A8, A9 and A10.

Figure A1
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Impulse-Response Functions, Australia

Source: Authors’ calculations.

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Figure A2
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Impulse-Response Functions, Canada

Source: Authors’ calculations.

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Figure A3
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Impulse-Response Functions, France

Source: Authors’ calculations.

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Figure A4
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Impulse-Response Functions, Japan

Source: Authors’ calculations.

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Figure A5
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Impulse-Response Functions, Sweden

Source: Authors’ calculations.

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Figure A6
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Impulse-Response Functions, United Kingdom

Source: Authors’ calculations.

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Figure A7
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Impulse-Response Functions for Alternate Ordering, Australia

Source: Authors’ calculations.

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Figure A8
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Impulse-Response Functions for Alternate Ordering, Canada

Source: Authors’ calculations.

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Figure A9
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Impulse-Response Functions for Alternate Ordering, France

Source: Authors’ calculations.

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Figure A10
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Impulse-Response Functions for Alternate Ordering, Japan

Source: Authors’ calculations.

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Figure A11
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Impulse-Response Functions for Alternate Ordering, Sweden

Source: Authors’ calculations.

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Figure A12
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Impulse-Response Functions for Alternate Ordering, United Kingdom

Source: Authors’ calculations.

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Bayoumi, T., Swiston, A. The Ties that Bind: Measuring International Bond Spillovers Using Inflation-Indexed Bond Yields. IMF Econ Rev 57, 366–406 (2010). https://doi.org/10.1057/imfsp.2009.13

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