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European Economic Review 119 (2019) 179–198
Contents lists available at ScienceDirect
European Economic Review
journal homepage: www.elsevier.com/locate/euroecorev
Increasing public debt and the role of central bank
independence for debt maturities
�
Lukas Nöh
Goethe University Frankfurt, RuW, Postbox 45, Theodor-W.-Adorno-Platz 4, Frankfurt 60629, Germany
a r t i c l e i n f o
Article history:
Received 3 August 2018
Accepted 23 July 2019
Available online 29 July 2019
JEL classification:
E 58
E 62
H 63
Keywords:
Sovereign debt
Debt management
Debt sustainability
Central bank independence
a b s t r a c t
Governments are interested in lengthening debt maturity to reduce fiscal, refinancing, and
default risks. On the other hand, investors may worry that highly indebted governments
will inflate away their debts when it is long-term. Based on a new panel dataset spanning
from 1990–2016 for 28 OECD countries, the paper finds that governments prolong the av-
erage debt maturity, if debt ratios increase. Interestingly, the marginal effect of the debt
ratio on average maturity is higher for countries with an independent central bank than
for countries with a more politically dependent one. There are two possible explanations
for this effect of central bank independence. First, an independent central bank reduces
the inflation risk premium that longer maturities otherwise would have. Second, govern-
ments with independent central banks are unable to use monetary policy for inflating and
reducing fiscal and interest rate shock burdens. Again, this calls for longer maturities to
reduce the relevant risks.
© 2019 Elsevier B.V. All rights reserved.
1. Introduction
Large debt is problematic for governments for several reasons. In the case of a macroeconomic shock, fiscal policy op-
tions are limited without increasing distortionary taxes. Furthermore, the government budget is more exposed to interest
rate hikes when numerous existing debts must be refinanced. A crisis of confidence in the ability to repay debts can lead to
liquidity problems that end in a default. For this reason, investors demand a premium when the debt is high. At the same
time, investors demand a premium for the possibility that the government lowers the value of its debt by high inflation.
Whereas debt reduction may be politically and economically difficult, debt management can limit the negative effects of
high debt with the choice of the maturity. A longer maturity can counteract the first three problems; however, it increases
investors’ inflation worries, because long-term debt increases the government’s incentive to increase inflation. As most cen-
tral banks have become more independent over the last two decades, governments have largely lost the means of inflating
away their debts. Therefore, debt management, in more advanced economies, has evolved towards downplaying inflation
risk concerns and instead lengthening debt maturity in order to reduce the fiscal, refinancing and default risks.
A positive relation between the maturity and the debt ratio is compatible with several arguments in the literature to
mitigate the problems of high debt. Cochrane (2001) , Angeletos (2002) and Buera and Nicolini (2004) argue for longer
� This paper is part of the research program of the Center for ”Sustainable Architecture for Finance in Europe” (SAFE). I thank three anonymous referee
for helpful comments. I am also grateful to Alfons Weichenrieder and all members of the Chair of Public Finance at Goethe University Frankfurt and to
discussants and participants of the 51st Annual Conference of the Canadian Economics Association 2017.
E-mail address: [email protected]
https://doi.org/10.1016/j.euroecorev.2019.07.007
0014-2921/© 2019 Elsevier B.V. All rights reserved.
180 L. Nöh / European Economic Review 119 (2019) 179–198
maturities to hedge against fiscal shocks. Following a macroeconomic shock that lowers government spending, higher inter-
est rates reduce the value of long-term debt. The reduction of refinancing risks due to interest rate fluctuations by means
of long-term debt is discussed, for example, by Barro (1995) , Bohn (1990) and Greenwood et al. (2015) . Long-term debt
reduces the exposure of the government budget to interest rate variations and hence lowers tax distortions. Finally, Alesina
et al. (1990) and Cole and Kehoe (1996, 20 0 0) show that, with too short maturity, a confidence crisis may arise and the
government may be forced to default. A long (and balanced) maturity structure helps to prevent a ( Diamond and Dybvig,
1983 ) liquidity crisis in which investors refuse to roll over the debt because they are afraid that everyone else will do the
same.
In Falcetti and Missale (2002) , the value of long-term debt is more sensitive to unexpected inflation, thus reducing the
need to increase distortionary taxes in the face of a macroeconomic shock. Moreover, they argue that delegating monetary
policy to an independent central bank is more efficient to reduce investors’ inflation concerns compared to using indexed,
foreign-currency or short-term debt. Using cross section analyses they find support for their theory in the lengthening of
debt maturity during the late 1980s until the mid 1990s with simultaneously increasing central bank independence.
At first glance, a positive relationship contrasts with the findings of Missale and Blanchard (1994) , who find a negative
relationship caused by inflation risk for the period 1960–1990. Their argument is based on a premium, which investors
demand when debt rises and inflation risks are substantial. The incentive to inflate debt is higher for governments facing
high average maturities because investors are tied for a longer period to the interest rate, without adjustment opportunities.
Hence, debt management can reduce interest rate costs by lowering average maturity, signaling that high inflation is not
intended and the inflation risk premium is not necessary.
This paper empirically analyzes the reaction of public debt management on changes in debt-to-GDP ratios over the last
27 years (1990–2016). Focusing on 28 advanced economies, it makes two contributions. First, the paper finds evidence of a
positive relationship between the average maturity of public debt and the debt-to-GDP ratio. Second, the paper shows that
the marginal effect of the debt-to-GDP ratio on average maturity is higher for countries with an independent central bank.
The positive effect of central bank independence on the correlation between maturity and debt ratio indicates that the
change in debt management policy is due to higher central bank independence. The inflation risk argument becomes less
relevant because the inflation risk decreased with the greater independence of central banks in recent decades. In addition,
the inability of governments with independent central banks to inflate away their debt may increase the exposure to in-
terest rate and default risks. Conversely, this suggests that countries which have a low credibility with respect to inflation
cannot reap the full benefits of longer maturities. Therefore, the empirical results can be interpreted to not only support the
arguments in favor of long-term debt but also to be compatible with the inflation risk arguments.
Two innovations contribute to empirical research on debt management in this paper. First, a novel dataset is compiled
on the average debt maturity of 28 OECD countries for the period 1990–2016. 1 The main data sources are publications
from national debt management agencies. In case of compatibility, these data are matched with the OECD dataset and
the empirical study by Missale (1999) . The resulting dataset allows a cross-country analysis of maturity developments. The
second innovation is the application of a dynamic panel estimation method in the field of debt management. To study the
reaction of debt management on debt-to-GDP ratios, the paper uses the pooled mean group estimator proposed by Pesaran
et al. (1999) . This method simultaneously allows homogeneous long-run relations and heterogeneous short-run adjustments.
The remainder of this paper is organized as follows: Section 2 provides the theoretical foundation for the research ques-
tion about the reaction of debt management to debt-to-GDP ratios. Section 3 describes the data, the empirical strategy, and
the method used for the pooled mean group estimator. Section 4 presents and discusses the empirical results as well as
several robustness checks. Finally, Section 5 concludes.
2. Developments in the relation between maturity and debt-to-GDP ratio
2.1. Empirical evidence
Empirical evidence for the relation between maturity and debt-to-GDP ratio can be found in Missale and Blanchard
(1994) . They describe a negative relation between both variables for the highly indebted countries Belgium, Ireland, and
Italy. Moreover they argue that higher debt ratios lead to higher incentives to inflate the debt away. In this situation, shorter
maturities underline the intention not to inflate because rewards from inflation are small with short maturities. Investors
can understand the short maturities as a signal of no inflation intentions, thus reducing their inflation risk and their required
risk premia.
Additionally, De Haan et al. (1995) find a negative relation for the Netherlands and Spain as well as some countries with
lower debt, such as Germany and the United Kingdom. For the US and Canada they find a positive relation which they
consider supportive of the confidence crisis theory of Alesina et al. (1990) .
However, these results were applicable prior to 1990. Considering new data from the 1990–2016 period, there is evidence
of a reversal of the negative relation for several countries. Fig. 1 depicts the average maturity and debt-to-GDP ratios for
28 OECD countries. The data show that, except for Hungary and Poland, each country has a higher average maturity in
1 In fact, Malta is the only non-OECD country in the dataset. However, for the sake of simplicity, the paper refers to the dataset as OECD countries.
L. Nöh / European Economic Review 119 (2019) 179–198 181
1020
3040
45
67
8
1990
1995
2000
2005
2010
2015
Australia
5060
7080
90
56
78
9
1990
1995
2000
2005
2010
2015
Austria
8010
012
014
0
24
68
1990
1995
2000
2005
2010
2015
Belgium
6080
100
45
67
1990
1995
2000
2005
2010
2015
Canada
1020
3040
50
12
34
56
7
1990
1995
2000
2005
2010
2015
Czech Republic
2050
80
35
79
11
1990
1995
2000
2005
2010
2015
Denmark
1030
5070
34
56
1990
1995
2000
2005
2010
2015
Finland
4060
8010
0
5.5
66.
57
7.5
1990
1995
2000
2005
2010
2015
France
4050
6070
80
45
67
1990
1995
2000
2005
2010
2015
Germany
7010
013
016
019
0
510
1520
1990
1995
2000
2005
2010
2015
Greece
5060
7080
90
34
56
7
1990
1995
2000
2005
2010
2015
Hungary
2040
6080
100
34
56
7
1990
1995
2000
2005
2010
2015
Iceland
2040
6080
1001
20
46
810
12
1990
1995
2000
2005
2010
2015
Ireland
6080
100
120
140
67
8
1990
1995
2000
2005
2010
2015
Israel
9011
013
0
23
45
67
1990
1995
2000
2005
2010
2015
Italy
5010
015
020
025
0
45
67
89
1990
1995
2000
2005
2010
2015
Japan
3050
7090
67
89
10
1990
1995
2000
2005
2010
2015
Malta
4050
6070
80
56
7
1990
1995
2000
2005
2010
2015
Netherlands
2030
4050
60
34
56
7
1990
1995
2000
2005
2010
2015
New Zealand
4060
8010
0120
140
34
56
7
1990
1995
2000
2005
2010
2015
Portugal
3040
5060
34
56
1990
1995
2000
2005
2010
2015
Poland
2030
4050
60
02
46
8
1990
1995
2000
2005
2010
2015
Slovakia
1030
5070
90
56
78
1990
1995
2000
2005
2010
2015
Slovenia30
6090
120
02
46
8
1990
1995
2000
2005
2010
2015
Spain
3040
5060
70
34
56
1990
1995
2000
2005
2010
2015
Sweden
3040
5060
67
89
10
1990
1995
2000
2005
2010
2015
Switzerland
2040
6080
100
810
1214
1618
1990
1995
2000
2005
2010
2015
United Kingdom
4060
8010
012
0
45
6
1990
1995
2000
2005
2010
2015
United States
Fig. 1. Average Maturity and Debt-to-GDP Ratio, 1990–2016.
Solid line (—) and left axis = Average Maturity in years. Dashed line ( −−) and right axis = Debt-to-GDP Ratio in percent. Source: Data from national
authorities, the OECD, and own calculations.
182 L. Nöh / European Economic Review 119 (2019) 179–198
2016 compared to 1990 (or the respective first year of available data). This contrasts with the period 1960–1990 reported
by Missale (1999) , when, for example, nine countries (Australia, Belgium, Canada, Ireland, Italy, Netherlands, Spain, Sweden,
and the UK) had sharp reductions of their maturities. In addition to the average maturity, the debt-to-GDP ratio also has
increased, albeit with fluctuations, over the last 27 years in most countries or has returned to almost the same level. As
a result, the overall relation appears to be positive for most countries. The simultaneous increase of maturity and debt-
to-GDP ratio suggests that the effects of inflation risk becomes less substantial compared to fiscal, refinancing, and default
risks.
2.2. Arguments for less inflation risk
Inflation risk was important for creditors until the mid 1980s, when inflation was both high and volatile. Since then, two
important changes have affected debt management. First, in the 1980s monetary policy changed, which shows up primarily
in lower inflation and reduced inflation volatility. Bernanke (2012) calls this period ‘The Great Moderation’ . Because low
inflation policies turned out to be sustainable, by the early 1990s the anti-inflation reputations of numerous governments
increased and subsequently the threat of debt inflating policies diminished. Hence, a signal against these policies through
shorter term maturities became less important.
The second change affecting debt management is the increase in central bank independence. This holds especially for
Europe, where many countries joined the Economic and Monetary Union (EMU) with a common central bank. In 1990 a core
group of countries, in what was then called the European Economic Community, the precursor of the European Union (EU),
began the first stages of an integrated economic and monetary system. The exchange rate mechanism connected European
currencies and inflation targets forced national central banks to coordinate their policies. Therefore, the increase in central
bank independence from national governments started already before the launch of the euro in 1999, extending even to
countries outside the euro area. Analyses measuring central bank independence by Garriga (2016) shows that some euro area
countries reformed their central banks to make them more independent before implementing the euro, namely Belgium,
France, Italy (in 1993), and Spain (in 1994). Because the influence of individual governments on inflation was reduced ( Klomp
and Haan, 2010 ), a signaling reaction of lower average maturities on rising debt ratios was no longer deemed necessary. 2
2.3. Arguments for long maturities
In view of rising public debts in recent decades, risks for the government budget become more important. Fig. 1 depicts
increasing debt-to-GDP ratios for the period 1990–2016 for most of the 28 countries analysed. Only Belgium, Denmark,
Israel, Netherlands and New Zealand have a substantial decline in their debt-to-GDP ratio. However, during this period
Belgium always had a debt-to-GDP ratio of at least 90% and, since the onset of the crisis, it has once again risen to above
100%. By 2016 Sweden, Hungary, and Ireland, after various fluctuations, once again have similar debt levels to those of 1990.
Nevertheless, the debt ratio of the latter two countries (and the Netherlands despite the decline) is still higher than the
60% allowed by the Maastricht criteria for sustainable debt. These findings show a trend of high debt-to-GDP ratios in most
countries which increases the burden of fiscal and interest rate shocks and heightens the possibility of an confidence crisis.
A positive relation between maturity and debt-to-GDP ratio is well documented for the US in Krishnamurthy and Vissing-
Jorgensen (2012) , Greenwood and Vayanos (2014) and Greenwood et al. (2015) . The latter show in their model that higher
debt leads to more fiscal risks as changes in interest rates have a greater impact on interest costs. These risks have negative
welfare effects due to greater volatility in tax rates and reductions in government investment programs. Thus, governments
are interested in insulation from interest rate fluctuations which can be reduced with long-term maturities. 3
Several events in the last 25 years have increased the risk of possible sovereign defaults. The crisis in the European Mon-
etary System in 1992 raised concerns about debt sustainability in many European countries. Spain and Italy, in particular,
experienced difficulties with rapidly rising risk premia ( Gros, 2014 ). The European sovereign debt crisis in 2010 showed
that even developed countries can come dangerously close to a possible default or, in the case of Greece, have to declare
a default. Alesina et al. (1990) argue that a self-fulfilling confidence crisis can be the consequence of investors’ anticipation
of excessive debt even before it actually reaches unsustainable levels. Long-term debt reduces the possibility of roll-over
problems in a confidence crisis and prevents liquidity problems which could lead to a partial default. Additionally, widely
discussed in the literature is a situation in which the government has to choose between bearing the welfare costs (e.g.
through distortionary taxes) of a shock or to risk default and accept the subsequent costs such as a temporary exclusion
from credit markets ( Mendoza and Yue, 2012 ). Investors anticipate this decision process and consequently demand higher
2 Before 1990, policies which resulted in debt inflation were an important issue for debt management. Calvo and Guidotti (1990) find optimal maturity
profiles in response to inflation risks depending on the precommitment possibilities of the government on future policy. Bohn (1991) finds a strong increase
in inflation incentives in the USA in the 1980s due to higher external nominal debt. More recent research does not attach much importance to the inflation
argument. Aizenman and Marion (2011) see some similarities for the US between inflation risks in the post World War II period and nowadays. But they
weaken this concern by arguing that in a globalized economy, foreign direct investors can easily move their activities to other countries in case of inflation
uncertainty. This pressure reduces the incentive to inflate away the debt. 3 Longer maturities with high levels of redemption concentrated at one time do not solve the whole problem. An evenly distributed redemption profile
over a long horizon, made possible by the extension of maturities, is also important.
L. Nöh / European Economic Review 119 (2019) 179–198 183
risk premia when debt is high. Long-term debt reduces the impact of high debt on fiscal risks, thereby lowering the incen-
tive to default.
Furthermore, the independence of central banks can play an important role in the attempt to reduce the problems of high
debt ratios. Although central bank independence lowers inflation risk, it also limits the government’s ability to intervene
with monetary instruments against interest rate shocks and confidence crisis. De Grauwe (2012) points out that members
of a monetary union lose their option of monetary financing for their budget. An independent central bank increases the
probability of a confidence crisis and reduces the possibilities to lower the burden of an interest rate shock. A no-bail-out
clause, as it is implemented in the EMU, strengthens this effect. On the other hand, Giordano and Tommasino (2011) refer
to evidence that more independence of a central bank leads to lower budget deficits and more sustainable debt. However,
their argument is precisely that the lack of support from the central bank in the event of a debt crisis, is what forces the
government to avoid the costs of a crisis for voters by taking on more sustainable debt. Debt management can partly tackle
this problem by lengthening maturities to insulate against interest rate shocks and make liquidity problems in a debt crisis
less likely. Therefore, the relationship between maturity and debt ratio depends on central bank independence.
3. Data and empirical strategy
3.1. Dataset
The data for average maturity consists mainly of national debt management information. Table 3 in Appendix A shows
details on the dataset. The most commonly used measure for maturity is the average time to maturity. Only Germany
publishes the slightly different average term to refixing, which in this case is comparable to average term to maturity.
The average time to refixing is defined by the weighted average time until all payments’ interest rates are renewed. The
Macaulay duration measures the time until the bond is repaid by its own internal cash flows. For the average time to
maturity, the weighted average time until all principal payments have to be redeemed is measured. 4 If available, the paper
uses the average maturity on marketable debt. For all other countries, the average maturity on total debt is used, which
in most cases develops like marketable debt or is equivalent to marketable debt. Missale and Blanchard (1994) argue that
the negative relation holds between debt-to-GDP ratio and effective maturity. Appendix A.2 shows that effective maturity
develops similar to maturity on total and marketable debt.
The following databases are used for explanatory variables. The long interest rate is the Government Bond Yield, obtained
from the OECD’s Financial Statistics Dataset. Inflation data is from the OECD Consumer Prices (Annual Inflation) measure.
The debt-to-GDP ratio is the general government gross debt taken from the IMF World Economic Outlook Database. From
the same source, the data on GDP growth is used. Data on debt held by non-residents and foreign currency debt is from
the OECD Central Government Debt Database and national debt management authorities. To determine the effect of central
bank independence, the paper uses the index introduced by Garriga (2016) 5 , a temporal expansion of the most commonly
used index developed by Cukierman et al. (1992) . It contains 16 components of independence for the 4 issues including:
CEO characteristics, policy formulation attributes, central bank objectives, and central bank limitations on lending to the
government. The index measures de jure, not de facto, independence, which is more important for the expectations of
investors. However, de facto independence is difficult to identify for investors and a long-term index is lacking. Furthermore,
this index includes only the period until 2012. Since there have not been any major central bank reforms in the countries
in the dataset in recent years, the index is assumed to be unchanged until 2016.
3.2. Empirical strategy
To analyze the relation between average maturity and debt-to-GDP ratio a panel error correction model with the pooled
mean group estimator (PMG) developed by Pesaran et al. (1999) is estimated. This estimation is commonly used in the em-
pirical literature for economic growth and fiscal policy (see, for example, Gemmell et al. (2011) and Arnold et al. (2011) for
OECD countries, and Unger (2017) for euro area countries.)
Basis for the error correction model is a ARDL(p,q) specification in which p is the number of lags for the dependent and
q for independent variables, and which reads
ma it =
p ∑
j=1
λi j ma i,t− j +
q ∑
j=0
β′ i j X i,t− j + μi + αit t + u it (1)
where i and t represent the country and time period, respectively. ma it is the log of average maturity and X it is a vector of
explanatory variables. μi is the intercept and represents a country-specific fixed effect, αit t indicates a country-specific time
trend and u i,t is a i.i.d. error term. Using the advantage of heterogeneous short-run dynamics for each country, the Schwarz
Bayesian Criterion (SBC) is applied to choose the lag order for each country separately ( Pesaran et al., 1999 ). Based on these
4 See OECD (2016) for definitions and calculations. 5 Data is available on https://dataverse.harvard.edu/dataset.xhtml?persistentId=doi:10.7910/DVN/I2BUGZ .
184 L. Nöh / European Economic Review 119 (2019) 179–198
criteria, an ARDL model with the maximum lag length of one is chosen. The robustness checks include tests for different
lag orders.
The explanatory variables in X consist of dr as the log of debt-to-GDP ratio and inf as inflation, which is not in log due to
negative inflation in many countries in recent years. Inflation serves as a variable for monetary policy variations. Although
inflation is included in the model, it does not fully control inflation risk, because the risk depends mainly on expected, not
on current inflation.
Debt-to-GDP ratio and inflation are the controls in the specification by Missale and Blanchard (1994) . A possible omitted
variable in their model is the interest rate. Because the reduction of interest rate costs is a key goal for debt management,
the adjustment of the maturity profile reacts to interest rate changes. Considering the usually increasing yield curve, an
increase in long-term interest rates raises the cost of higher average maturity. 6 As there are some examples of a decreas-
ing yield curve ( Alfaro and Kanczuk, 2009 ) which also appear for some years in the dataset of this paper, several control
variables for the interest rate ir are included. Whereas the 10-year government bond rate is the traditional measure for the
long-term interest rate, for robustness the monetary policy rate is also included for central bank influence on the interest
rate and the difference between both in order to capture the slope of the yield curve. Additionally, GDP growth gr serves as
a control for economic development.
To analyze the effect of independent central banks, the measure for central bank independence and an interaction term
is included. The variable cbi is the measure for central bank independence with a range between 0 (totally dependent) and
1 (totally independent). Finally, cbiit is an interaction term and multiplies the indicator for central bank independence with
the log-debt-to-GDP ratio variable. The interaction term measures the effect of central bank independence on the relation
between debt-to-GDP ratio and average maturity.
Following Pesaran et al. (2001) , the ARDL model can include I(0) and I(1) variables simultaneously, as long as no variable
exceeds an I(1) process. Considering the time series for each country suggests a non-stationary dataset. Various unit root
tests in Table 4 in Appendix B indicate a mixture of I(0) and I(1) variables.
Pesaran et al. (1999) show that the pooled mean group estimator based on an error-correction model does not need
pretesting for cointegration. The model-included error-correction term tests for long-run relationships and finds a significant
long-run relationship. As it is common in the PMG literature to run additional tests for cointegration, the paper follows
( Unger, 2017 ) and applies the Kao and the Pedroni methods. Table 5 in Appendix B shows the results for the Kao Augmented
Dicky–Fuller (ADF), the Pedroni Phillips–Perron, and the Pedroni ADF tests. The last two tests include panel and group
mean statistics. For all specifications and tests, the null hypotheses of no cointegration is rejected. Hence, together with the
unit root test results, the analysis proceeds with the assumption that the variables are non-stationary (in differences) and
cointegrated.
The validity of this estimator is compared to the mean group (MG) estimator ( Pesaran and Smith, 1995 ) and the con-
ventional dynamic fixed effects (DFE) estimator. The MG estimator assumes heterogeneity in the long- and short-run and
pools the results of each country’s regression. The DFE estimator assumes homogeneity in the long- and short-run and only
allows the intercept to differ. While the MG estimator produces consistent estimates, it is inefficient in case of homogeneity
in the long-run. As Pesaran et al. (1999) argue, the MG is highly sensitive to outlier countries. On the other hand, the PMG
estimator is rather robust to outliers. The DFE estimator is efficient, but can be inconsistent in case of failure of the ho-
mogeneity assumption. The PMG estimator is efficient, if the homogeneity assumption holds in the long-run. The Hausman
test helps to determine which estimator is preferred. Under the null hypothesis the coefficients do not differ systematically
between the models. Because none of the results can reject the null hypothesis, the PMG and its homogeneity assumption
is consistent for all model specifications. Therefore, the PMG is selected because it is the most efficient estimator.
The PMG estimator assumes the long-run dynamics to be homogeneous and the short-run dynamics to be heterogeneous
across countries. Heterogeneity in the short run is assumed to be reasonable because bond issue strategies are usually
scheduled approximately every 12 months, depending on the debt management policy. Adjustments to crises are made
independently in each country, depending on the length and intensity of any shocks. Depending on strategies and issuing
frequencies, countries are heterogeneous in their short-term reactions to macroeconomic changes. In the long-run, however,
debt management adaptions to macroeconomic and monetary changes as well as common risk assessments are plausible. It
is unlikely that debt management across highly integrated OECD countries have vastly different estimates for inflation and
fiscal risk developments.
The error-correction model for the ARDL specification in equation (1) reads
�ma i,t = φi [ ma i,t−1 − θ′ i 1 X i,t−1 ] +
p−1 ∑
j=1
λi j �ma i,t− j +
q −1 ∑
j=0
δ′ i j �X i,t− j + μi + αit t + εi,t (2)
where λ and δ are the short-run coefficients for the dependent and independent variables, respectively. The vector θ ′ rep-
resents the long-run coefficients of the independent variables being analyzed. φ is the error-correcting speed of adjustment
term. If this term equals zero, there would be no long-run relation. Otherwise, φi < 0, implies a co-integration long-run
relation between ma it and X it .
6 However, the theory for an increasing yield curve is controversial, see, for example, ( Barro, 1997 ).
L. Nöh / European Economic Review 119 (2019) 179–198 185
Table 1
Relation between debt-to-GDP ratio and average maturity.
Variables Baseline h Incl. CBI h Incl. CBIIT h
Long-run coefficients
Debt-to-GDP ratio 0.162 ∗∗∗ 0.62 0.192 ∗∗∗ 0.43 0.055 0.14
(0.028) (0.43) (0.031) (0.51) (0.042) (0.70)
Inflation −0.001 0.83 −0.009 ∗∗ 0.02 0.000 0.54
(0.004) (0.36) (0.005) (0.88) (0.004) (0.46)
GDP growth rate −0.019 ∗∗∗ 0.86 −0.016 ∗∗∗ 0.81 −0.018 ∗∗∗ 1.03
(0.004) (0.35) (0.004) (0.37) (0.003) (0.31)
Long-term rate −0.025 ∗∗∗ 1.57 −0.038 ∗∗∗ 0.49 −0.032 ∗∗∗ 0.91
(0.005) (0.21) (0.005) (0.48) (0.004) (0.34)
CBI 0.196 ∗∗∗ 1.00 −0.805 ∗∗∗ 0.36
(0.059) (0.32) (0.234) (0.55)
CBIIT 0.226 ∗∗∗ 0.39
(0.055) (0.53)
Error correction term −0.474 ∗∗∗ −0.458 ∗∗∗ −0.525 ∗∗∗
(0.066) (0.068) (0.074)
Countries 28 28 28
Observations 641 641 641
Notes: Pooled mean group estimates for log of average maturity including controls for country
and time trend effects. Debt-to-GDP ratio is in log. CBI is a measure of central bank indepen-
dence. CBIIT is the interaction term between central bank independence and debt-to-GDP ra-
tio. Robust SE are in parentheses for the coefficients. Significance is indicated by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1. The Hausman test rejects significant differences to the MG estimator.
The Hausman test h rejects significant differences to the MG estimator. For the Hausman test,
p -values are in parentheses.
To deal with the common factor problem, the data is cross-section demeaned, as Pesaran et al. (1999) propose. With this
procedure, it is assumed that errors are independent between groups.
4. Estimation results
4.1. Positive effect of debt-to-GDP ratio on average maturity
Table 1 reports the results for the relation between debt-to-GDP ratio and average maturity of the dynamic panel re-
gression. The upper part of the table shows the long-run coefficients. 7 The error correction coefficient in the lower part is
negative and statistically significant for all estimators. Together with the requirement to be less than one in absolute value,
this result indicates evidence for (cointegrating) long-run relations. Thus, the coefficient confirms the long-term relationship
described in the previous section. The Hausman tests and the corresponding p-values indicate that the null hypothesis with
the long-run homogeneity assumption is not rejected, and the PMG estimator is consistent for all variables of interest.
The result in Column 1 shows strong statistical significance for a positive effect of debt-to-GDP ratio on average maturity.
According to the coefficient, a one percent increase of the debt-to-GDP ratio leads to an increase of the average maturity
of 0.162%. This result is consistent with the arguments which demand higher maturities against the negative effects of high
debt-to-GDP ratios on debt sustainability. It goes against the theory that a shorter maturity is needed, as the debt increases,
to commit to low inflation. The changes in maturity structure reflect strong shifts in the bond markets. An increase in the
government debt ratio from, for example, 60% to 66% leads the average maturity to rise from 6 to about 6.1 years. An
increase in the average maturity of even 0.1 years is only possible with even longer maturities of roll-over debt. If, as in
this example, on average one-sixth of the public debt is refinanced every year, this means an even stronger increase in the
supply of long-term government bonds. 8
The coefficient for inflation has no significant effect and the GDP growth rate has a negative effect on average maturity.
These results are also consistent with the two theories. Debt management has no reason to react on inflation, but smaller
GDP growth has a negative impact on tax revenues and raises the debt-to-GDP ratio. Accordingly, fiscal risks increase and
debt management lengthens maturities to mitigate these risks. The effect of the long-term interest rate is negative, which
may not be surprising, as higher long-term bond prices should cause debt management to use less long-term and more
short-term debt. The effect of debt on maturity is thus a significant factor in addition to long-term interest rates. The ro-
bustness checks deal more closely with the effects of interest rates on maturity. However, it should be noted that expected
inflation affects long-term interest rates. Thus, a negative effect of long-term interest rates may speak in part for the argu-
7 Short-run results are skipped as they provide no informational value for the purpose of this paper. 8 In an example of a country with sovereign debt of 120 billion euros, the 10% increase in debt is 12 billion euros. Assuming a flat maturity structure
with an average maturity of 6 years, 10 billion euros must be refinanced each year. The remaining 110 billion euros have a maturity of 5.5 years at the end
of the year. To increase the total average maturity to 6.1 years, the total amount of newly issued debt, 22 billion euros, needs a maturity of 9.1 years.
186 L. Nöh / European Economic Review 119 (2019) 179–198
Table 2
Subsamples for high and low debt-to-GDP ratio countries.
Variables Debt Ratio > 60% Debt Ratio < 60%
Baseline h Incl. CBIIT h Baseline h Incl. CBIIT h
Long-run coefficients
Debt-to-GDP ratio 0.538 ∗∗∗ 0.07 0.051 0.50 0.142 ∗∗∗ 0.18 0.073 ∗ 1.14
(0.080) (0.79) (0.073) (0.48) (0.032) (0.67) (0.045) (0.29)
Inflation −0.005 0.00 −0.010 ∗∗ 0.28 −0.029 ∗∗∗ 0.38 −0.025 ∗∗∗ 1.13
(0.008) (1.00) (0.005) (0.60) (0.005) (0.76) (0.005) (0.29)
GDP growth rate −0.010 ∗∗ 2.78 −0.008 ∗ 0.15 −0.024 ∗∗∗ 0.00 −0.020 ∗∗∗ 0.85
(0.005) (0.10) (0.004) (0.70) (0.006) (0.99) (0.006) (0.36)
Long-term rate 0.007 0.75 0.004 0.01 −0.035 ∗∗∗ 0.90 −0.047 ∗∗∗ 0.72
(0.007) (0.39) (0.005) (0.91) (0.006) (0.34) (0.008) (0.40)
CBI −1.582 ∗∗∗ 0.63 −1.083 ∗∗∗ 1.29
(0.439) (0.43) (0.248) (0.26)
CBIIT 0.346 ∗∗∗ 0.64 0.277 ∗∗∗ 1.28
(0.094) (0.42) (0.072) (0.26)
Error correction term −0.317 ∗∗∗ −0.403 ∗∗∗ −0.631 ∗∗∗ −0.549 ∗∗∗
(0.075) (0.080) (0.114) (0.120)
Countries 16 16 12 12
Observations 378 378 263 263
Notes: Pooled mean group estimates for log of average maturity including controls for country and time trend effects.
Debt-to-GDP ratio is in log. CBI is a measure of central bank independence. CBIIT is the interaction term between central
bank independence and debt-to-GDP ratio. Robust SE are in parentheses for the coefficients. Significance is indicated by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1. The Hausman test h rejects significant differences to the MG estimator. For the Hausman
test, p -values are in parentheses.
ment of Missale and Blanchard (1994) . Given that inflation and hence inflation expectations were comparatively low during
the period under consideration, this effect should be very small.
The effort s of debt management to minimize default risks and the related premium will be stronger, if the debt ratio
not only rises but is also at a higher level. Eichengreen et al. (2001) mention that safe countries are more willing to issue
short-term debt to lower interest rate costs, as default risk premia are of lesser relevance for low-debt countries. Table 2
shows the debt-to-GDP ratio coefficient for a subsample of high-debt countries compared to those with low debt-to-GDP
ratios. The high debt ratio sample includes countries with an average debt-to-GDP ratio larger than 60%. The countersample
includes countries with ratios lower than 60%. Although the statistical significance of such small samples should not be
overstated, the result gives an impression that the absolute level of debt ratios also has an influence. However, the effect of
the debt ratio in low debt-countries is smaller yet positive and significant. This suggests that even low-debt countries are
interested in reducing risks of fiscal and interest rate shocks, despite these risks being smaller with low debt ratios. In the
sovereign debt crisis, Spain has shown that even countries with a comparatively low debt level, for example due to a real
estate or banking crisis, can have very rapidly rising debts.
De Haan et al. (1995) argue that the positive relation holds only for periods with rising debt. The correlations in
Fig. 1 show a positive relation for declining debt ratios, among others, for Finland and Hungary. The correlation seems
to be negative for declining debt ratio periods in Belgium, Canada, and Italy. These countries have high debt ratios despite
they are declining during certain periods. In their situation, risks can still be so high, despite this positive development,
that a reduction of the maturity would only make sense with even lower debt ratios. For empirical analyses, the downturns
using annual data, are too short. However, a positive relationship only in periods of rising debt would not be a problem for
the reasoning in this paper. Due to the low risk of inflation, in times of stagnating or falling debt ratios, it would not be
necessary to immediately reduce the maturity. Instead, especially when high debt levels still prevail, fiscal, refinancing, and
default risks could be lowered further with longer maturities.
The results could raise endogeneity concerns. The long-term interest rate in particular could be driven by changes of the
average maturity. The aim of debt management is to reduce risks. As far as this reduces also risks of investors (e.g., default
risk) they lower their demand for risk premia and hence the long-term interest rate. While the argument of this paper
focuses on the effect of long-term rates on maturity changes, the causality is to some extent controversial. As the long-term
rate changes interest costs and hence debt-to-GDP ratio, even this variable could be effected by endogeneity problems. The
same holds true for inflation expectations which could be influenced by changes of maturity. Nevertheless, the starting point
seems to be the action of debt management based on the costs (interest rates) and the risks (debt ratio).
4.2. Influence of central bank independence
Column 2 of Table 1 indicates a significantly positive effect of central bank independence on average maturity. Reforming
a central bank so that is moves from being completely dependent to completely independent increases average maturity
by 19.6% ( CBI measure changes from 0 to 1). Although such an extreme case is highly abstract, the adjustments of the
L. Nöh / European Economic Review 119 (2019) 179–198 187
central bank system in the early 1990s to prepare for the EMU have led to a strong increase in central bank independence.
For example, as measured in France and Italy, independence has increased by 0.5. According to the coefficient in Column
2, such reforms lead to an almost 10% increase of average maturity. Inflation and the long-term interest rate again have
no effect on maturity; GDP growth has a negative effect. The effect of debt-to-GDP ratio is similar to the results found in
Column 1.
The eurozone has a special role to play in that it has created high levels of central bank independence for many countries
that previously relied exclusively on their own national central banks, which in most cases were less independent prior to
the launch of the euro. The change in independence for countries in the euro area, as well as other countries, some of which
implemented reforms of similar or even higher independence, is reflected in the measure of central bank independence. Half
of the countries in the dataset (14 out of 28) are now part of the eurozone, which naturally influences the results of the
estimates. In Appendix C, Table 7 , the subsamples are divided into Eurozone countries and those outside, thus revealing
the strong influence of the Eurozone. Especially in these euro area countries, the independence of monetary policy has
increased sharply, resulting in significantly lower inflation yet higher default risks. However, one should keep in mind for
this robustness check that these countries have had their own less independent central banks for a significant portion of
the considered period. Yet some countries outside the eurozone also have increased their central bank independence; for
example, the United Kingdom, which is also reflected in longer debt maturities. Nevertheless, in the subsample excluding
euro area countries, countries with relatively low central bank independence tend to dominate.
Column 3 of Table 1 includes the interaction term between central bank independence and debt-to-GDP ratio. The in-
clusion of an interaction term changes the interpretation of the underlying variables. The insignificant debt-to-GDP ratio
coefficient is the hypothetical effect for a country with a completely dependent central bank, cbi = 0 . This is an unreal-
istic scenario in the dataset, in which the average central bank independence is 0.63. In order to interpret the debt-to-
GDP ratio coefficient, the coefficient of the interaction term must also be considered. The coefficient for debt-to-GDP ratio
consists of the coefficients βdr and βcbiit in βdr + βcbi −it · cbi . For the coefficients in Column 3 and cbi = 0 . 63 it becomes:
0 . 055 + 0 . 226 · 0 . 63 . This results in a positive coefficient for debt-to-GDP ratio: 0.197, which is similar to the coefficient
in other specifications. The same holds for the interpretation of the central bank independence coefficient. For an average
debt-to-GDP ratio of 65%, the central bank independence coefficient is 0.137.
The positive and significant interaction term provides evidence for the hypothesis that the more independent the central
bank is, the stronger the relation between debt-to-GDP ratio and average maturity. The positive effect of central bank in-
dependence on this relation is consistent with the theory that this independence reduces debt management’s inflation risk
concerns and promotes a stronger focus on fiscal, refinancing, and default risks with longer maturities. Moreover, Column 3
of Table 1 depicts the significant negative effects of the long-term interest rate and the growth rate. This is in line with the
arguments for the observed effects in Column 1.
After the financial crisis, central banks began to buy large amounts of long-term government debt. The aim of, for ex-
ample, Operation Twist in the US or the ECB’s Public Sector Purchase Program is to reduce long-term interest rates. Effects
of these programs should not be included in the debt-to-GDP ratio coefficient, because the results in Table 1 control for
the long-term interest rate (and in the robustness checks for further interest rates). Further, this should not be the case, as
the programs were only introduced late into the observation period; Operation Twist started in 2011 and the ECB program
mainly in 2015. Nevertheless, a robustness check in Table 6 in Appendix C shows, that also for the period 1990–2010 the
debt-to-GDP ratio coefficient is positive and significant. Unsurprisingly, the coefficient is smaller than for the entire dataset,
as the post-crisis period is characterized by rising default risks and corresponding maturity extensions. In most countries,
this has been achieved without central banks buying debt by mid-2015.
4.3. Foreign debt
4.3.1. Foreign currency debt
In addition to the choice of maturity structure, debt management can also influence the interest costs with the currency
of the issued bond. Issuing debt in foreign currencies may reduce inflation risk premia, but as Falcetti and Missale (2002) ar-
gue, an independent central bank is more effective in reducing inflation risks. Their arguments for more nominal and larger
maturity debt together with their empirical findings for the beginning of the 1990s, finds support in the results of this pa-
per. Fig. 2 shows that only a few countries in the dataset have significant shares of foreign currency debt. Large countries
such as Germany, Japan, the UK or the US had, at least since 1990, no or almost no foreign currency debt. In addition, the
shares in foreign currency bonds of those countries that had larger shares at the beginning of the 1990s fell significantly
over the last 20 years.
Only Poland, Hungary, Iceland, Sweden, Israel, and the Czech Republic have more than 10% of foreign currency debt.
For these countries, the ratio of the average maturity between total debt and foreign currency debt varies widely. Fig. 4 in
Appendix A.3 shows the difference between the average maturity of total debt and foreign currency debt. For example,
whereas in Israel the maturity of foreign currency debt is always lower than the total debt, in Poland the opposite is the
case.
Due to the very low importance of foreign currency debt compared to total debt, it is very unlikely that they affect
the ratio between maturity and debt-to-GDP ratio. Rather, the decline in foreign currency debt supports the argument that
the increasing independence of central banks reduces inflation risks. Debt management needs less foreign currency debt to
188 L. Nöh / European Economic Review 119 (2019) 179–198
Fig. 2. Fraction of Foreign Currency Debt on Total Debt in %, 1990–2016.
Notes: Source: OECD and national debt management organizations.
reduce its inflation risk premia, just as it requires less short-term debt. With the extension of maturities and the reduction
of foreign currency bonds, the independence of the central bank continues what Falcetti and Missale (2002) have already
argued for the beginning of the 1990s.
4.3.2. Debt held by non-residents
The integration of global financial markets allowed many countries to find international investors for their debt. Access
to long-term debt depends, among other things, on whether access to the international bond market can be established.
However, a higher proportion of non-resident debt holders could increase the incentive to default. Reinhart and Rogoff
(2011) show that most defaults are linked to high shares of external debt. Changes in maturity may therefore be due to
changes in the share of external debt.
However, in OECD data no relationship between external debt and average maturity can be found. Fig. 5 in appendix
E shows that, for example Japan and United Kingdom, two countries with high and rising maturities, have low shares of
external debt. In addition, the regression in Table 11 also in Appendix E shows that the share of external debt has no effect
on average maturity. Debt in foreign ownership seems not to influence debt management policy. A reason for this result
could be that for advanced economies a default, and thus a poor valuation by the financial markets, causes great damage
for years. Even with small shares of external debt, with highly integrated financial markets the damage to the domestic
economy can be considerable. Governments of advanced economies try to avoid any appearance of default intentions in
order to not lose access to international markets. Furthermore, it is impossible to default exclusively on debt held by non-
residents, without also affecting residents.
4.4. Further robustness checks
Possible concerns can arise if there is too little variation in the central bank independence measure. Countries with no
variation in independence are not considered for independence variables due to the first differences method. 9 A robustness
check on this problem is conducted by testing the effect of debt-to-GDP ratio on average maturity for two sub-samples of
countries with high or low central bank independence. Table 8 in Appendix D depicts positive (larger than for the entire
9 This concerns four countries in the dataset, namely Denmark, Poland, Sweden, and the US.
L. Nöh / European Economic Review 119 (2019) 179–198 189
dataset) effect of the debt-to-GDP ratio on maturity for countries with a high level of central bank independence and a small,
not significance effect for those countries with lower independence rating. This result confirms previous results, including
the interaction term. The high independence group consists of countries with an average (over the entire observation period)
independence measure higher than 0.63 (the average over all countries). The comparative group includes countries with an
average independence measure below 0.63. However, statements on panel data sets with relatively small group sizes should
be treated with caution.
To further confirm the results, different lag orders are cross checked. Pesaran et al. (1999) emphasize that in an unbal-
anced panel with a small T, fixed lag order might show misspecifications. Nevertheless, it is common in the literature to use
a fixed lag order at least for robustness checks. Table 9 in Appendix D confirms the positive relation between debt-to-GDP
ratio and the average maturity in Column 1. Inflation again has no effect and GDP growth and long-term interest rate have
negative effects. Column 2 confirms these results, but shows no significant effect for central bank independence when us-
ing fixed lag orders. The interaction term in Column 3 is similar to the results with flexible lags and confirms the positive
influence of central bank independence on the maturity-debt relation. Again, the debt-to-GDP ratio coefficient is distorted
by the interaction term. In contrast to the flexible lag estimation, the inflation rate has a significant negative effect, whereas
the long-term interest rate has a significant positive effect.
To address possible endogeneity concerns, a maximum lag order of two can be implemented. This is only possible for at
most three variables, with zero lags for all other variables. The reason is that the number of parameters must be smaller
than the minimum number of periods for all groups. For several combinations of this specification, the results are robust.
This result holds especially for the main variables of interest, namely the debt-to-GDP ratio coefficient and the interaction
term. Nevertheless, the robustness would be improved further with more periods, which can be achieved with a longer time
horizon.
In recent years, great progress has been made in the literature on the effects of debt changes on interest rates. Beetsma
et al. (2017) and Krishnamurthy and Vissing-Jorgensen (2012) find evidence for such effects. Hence, a reversed causality
problem for the interest rate control variable can emerge. Furthermore, a general decrease in interest rates can be observed
in almost all advanced economies. Lower inflation and globalization of financial markets with more foreign creditors are two
main explanations. These influences on the prices of government debt also affect the development of maturities. However,
once controlling for the interest rate, the effect of debt-to-GDP ratio on average maturity should not be distorted. To confirm
this, Table 10 in Appendix D shows estimations for different interest rate measures. The overall result is that the debt-to-
GDP ratio and the interaction term are still positive and significant. However, the coefficients for the interest rate change in
several specifications. The monetary policy rate is used as a proxy for short-term interest rates in Column 1 and it shows
a significant negative coefficient, but a smaller and less significant effect for the long-term interest rate compared to the
main results. That could be interpreted to mean that debt management relies to some extend on changes in central bank
policy and not only on market interest rates on long-term debt. Combining both measures to an interest rate spread depicts
a positive effect on average maturity for Columns 2 and 3, but a negative effect in Column 4. An increasing spread between
long- and short-term rates signals increasing interest rates in the future and hence can force debt management to conserve
lower rates today. On the contrary, a rising spread increases the price of long-term debt compared to short-term debt and
can lead debt management to shorten average maturity. A collinearity problem between inflation and the long-term interest
rate seems possible at first glance. Exclusion of inflation in the regressions doubles the interest rate coefficient. The exclusion
of the interest rate leads to an significantly positive inflation coefficient. Both cases do not crucially change the coefficients
of interest, debt-to-GDP ratio or the interaction term. For these reasons, a collinearity problem is not an important factor in
the estimation.
5. Concluding remarks
This paper finds evidence for a statistically significant and positive impact of the debt-to-GDP ratios on average maturity.
This result revises most empirical findings in the past. The reason for this positive correlation can be explained by the second
result of the paper: The marginal effect of the debt ratio on average maturity is higher for countries with an independent
central bank than for countries with a more politically dependent one. There are two possible explanations for this effect
of central bank independence. First, an independent central bank reduces the inflation risk for investors such that their risk
premia can decrease although governments issue longer-term bonds. The government has less influence on monetary policy
and therefore cannot take advantage of long maturities with high inflation. Second, it urges debt management to pay more
attention budget risks. The inability of governments with independent central banks to use monetary policy removes the
possibility to inflate away their debt, which increase the probability of default. In addition, a monetary policy intervention
to mitigate fiscal and interest rate shocks becomes less likely.
These findings, obtained using a panel dataset spanning from 1990–2016 for 28 OECD countries, suggest that debt man-
agement has changed its policy concerning risk assessment. Greater central bank independence is a reason that debt man-
agement focuses less on inflation risks and more on fiscal, refinancing and default risk. With this revaluation, the average
maturity period has lengthened. The empirical finding contrasts with Missale and Blanchard (1994) , who find a negative
relation for the period 1960–1990. However, because their results are valid during a time of high inflation risks, a reversal
of the correlation after increased central bank independence is in line with their arguments. There are many reasons to
190 L. Nöh / European Economic Review 119 (2019) 179–198
increase maturities as discussed in this paper. However, this paper argues that higher central bank independence favors
debt management interest in longer maturities.
These findings have important implications for the relationship between debt management and monetary policy. Coun-
tries with an independent central bank generally seek to prevent a confidence crisis by using higher interest rates for long-
term debt. These costs could be reduced if the central bank is allowed to intervene as a lender of last resort in the event of
a crisis. However, this concession would come at the expense of a less independent central bank, potentially raising inflation
risks. For countries with a less independent central bank, especially in emerging countries, greater independence could help
to gain better access to the long-term bond market and thereby build a stronger debt management strategy.
Appendix A. Dataset and maturity measures
A1. Dataset
Table 3 reports the published type (Columns 2–4) of maturity measure and the period (Column 5) of available data
from the debt management authority. Debt management data are combined with two additional sources. The first source
is the OECD Central Government Debt Database, which only records and reports on the composition of government debt
until 2010. Furthermore, this database does not include many countries and has numerous missing entries. The second
source is the dataset collected by Missale (1999) until 1995. For Malta and the Netherlands data from the ECB is added.
In general, national debt management data are prioritized and other sources are added only when deemed necessary to
increase comparability of the measure. Column 6 of Table 3 shows the entire period for which data could be used in the
analyses. For some countries, the time series is limited by the availability of long-term interest rates.
Table 3
Data on maturity measures.
Country Maturity Duration Repricing Debt management Total Base
Australia x x x 1996-2016 1990-2016 M
Austria x - - 1981-2016 1990-2016 T,M,I,C
Belgium x x - 1999-2016 1990-2016 T,C
Canada x - - 1955-2016 1990-2016 T
Czech Rep x x x 2002-2016 1998-2016 T,M
Denmark x x - 2002-2016 1990-2016 T,L
Finland x x x 1982-2016 1990-2016 M
France x - - 1997-2016 1990-2016 M,L
Germany - - x 1999-2016 1990-2016 M
Greece x - - 1999-2016 1998-2016 T
Hungary x - - 2008-2016 1997-2016 C ∗
Iceland x - - 2011-2016 1993-2016 T,I,C
Ireland x - - 1998-2016 1990-2016 L ∗
Israel x x - 1995-2016 1997-2016 T,C
Italy x x x 1982-2016 1990-2016 M
Japan x - - 1996-2016 1990-2016 T
Malta x - - 2009-2016 1999-2016 L
Netherlands x - - 2012-2016 1990-2016 L,C
New Zealand x - - 1999-2016 1992-2016 M,I
Portugal x x - 2001-2016 1996-2016 T,T ∗
Poland x x x 1997-2016 1997-2016 T,M,C,L
Slovakia x x - 2009-2016 1997-2016 T
Slovenia x x - 2001-2016 2000-2016 T,L
Spain x x - 2000-2016 1990-2016 T,I,C,CB
Sweden x x - 1998-2016 1990-2016 T,I,C
Switzerland x - - 2000-2016 2000-2016 M
UK x x - 2000-2016 1990-2016 T,L
US x - - 1980-2016 1990-2016 M
∗
Notes: Data collected from national debt management authorities. Maturity is the Average Term to Ma-
turity, Duration is the Modified or Macaulay Duration, and Repricing is the Average Term to Repricing.
Calculation base for maturity is T = total portfolio, M = marketable, I = by instruments, C = by currency,
L = long-term bonds and CB = central bank holdings. Hungary’s debt is calculated using the maturity pro-
file (C ∗). For Ireland EU/IMF program bonds are included (L ∗) and for Portugal those bonds are sepa-
rately considered in total (T ∗). New Zealand’s maturity is calculated from historical issuance tables. The
US publishes Marketable Interest-Bearing Public Debt Held by Private Investors (M
∗).
L. Nöh / European Economic Review 119 (2019) 179–198 191
A2. Effective maturity
A crucial point in the theory by Missale and Blanchard (1994) is the difference between conventional and effective ma-
turity. Because investors’ concerns about inflation is focused on fixed rate domestic currency debt, considering total or mar-
ketable debt does not sufficiently capture the problem. To capture the maturity effected by inflation, they introduce the
effective maturity . This variable neglects price-level indexed and foreign currency debt, because national inflation cannot
effect this kind of debt.
For two reasons marketable debt is also appropriate. First, as Fig. 3 depicts, the course of conventional and effective
maturity is similar, which is most important for comparisons with debt-to-GDP ratio. Second, De Haan et al. (1995) find for
the Missale dataset that changes between effective and conventional maturity occur only for Ireland and Italy. Data show
that on the whole, albeit with varied fluctuations, some of which result from special issuance programs (e.g., in Ireland
and Italy), both conventional and effective maturity declined in the pre-millennial decades. The share of nominal on total
debt strongly increased over the last 20 years and conventional and effective maturity aligned further. According to Missale
(1999) , this approximation already started in the late 1980s.
Fig. 3. Conventional and Effective Maturity
Notes: Solid (–) = Conventional maturity in years. Dashed ( −−) = Effective maturity in years. Data source: ( Missale, 1999 ).
192 L. Nöh / European Economic Review 119 (2019) 179–198
A3. Foreign currency debt
Fig. 4. Difference Between Average Maturity of Total and Foreign Debt, 1997–2016.
Notes: Source: OECD and national debt management organizations.
Appendix B. Unit root and cointegration tests
Table 4 shows the results of the first generation unit root tests by Im et al. (2003) and the Fisher test implemented by
Maddala and Wu (1999) . Both tests allow for an unbalanced dataset and heterogeneity between autoregressive coefficients.
Additionally, the second generation unit root test by Pesaran (2007) is applied. Compared to the first generation tests, the
second allows for cross-section dependencies, which can be an issue due to macroeconomic linkages and unobserved com-
mon factors among OECD economies. The results show that all three tests reject the null hypothesis of the presence of unit
roots in differences and are at least a I(1) process. For inflation, GDP growth, and the long-term interest rate, the tests give
evidence for an I(0) process.
L. Nöh / European Economic Review 119 (2019) 179–198 193
Table 4
Unit root test results.
Variables Level First Difference
IPS Fisher Pesaran IPS Fisher Pesaran
Average Maturity trend −0.63 47.64 2.42 −6.94 196.45 −4.35
(0.73) (0.65) (0.99) (0.00) (0.00) (0.00)
no trend −1.85 59.67 −1.30 −8.60 209.68 −5.03
(0.03) (0.22) (0.09) (0.00) (0.00) (0.00)
Debt-to-GDP ratio trend −0.91 63.09 0.05 −4.62 105.93 −2.57
(0.18) (0.14) (0.52) (0.00) (0.00) (0.00)
no trend −1.82 93.59 −1.29 −7.12 157.75 −4.07
(0.03) (0.00) (0.09) (0.00) (0.00) (0.00)
Inflation trend −6.59 194.72 −3.34 −14.60 380.61 −9.52
(0.00) (0.00) (0.00) (0.00) (0.00) (0.00)
no trend −6.44 243.26 −5.68 −17.05 501.34 −11.56
(0.00) (0.00) (0.00) (0.00) (0.00) (0.00)
GDP growth trend −6.95 213.89 −4.40 −14.04 466.95 −9.04
(0.00) (0.00) (0.00) (0.00) (0.00) (0.00)
no trend −10.04 255.84 −7.68 −17.25 608.69 −11.66
(0.00) (0.00) (0.00) (0.00) (0.00) (0.00)
Long-term rate trend −3.05 105.53 −3.32 −10.24 224.55 −7.77
(0.00) (0.00) (0.00) (0.00) (0.00) (0.00)
no trend −4.57 51.61 −1.22 −12.92 312.36 −9.31
(0.00) (0.49) (0.11) (0.00) (0.00) (0.00)
cbi trend −0.853 26.81 1.56 −10.27 165.37 −2.054
(0.20) (0.99) (0.94) (0.00) (0.00) (0.02)
no trend −5.25 68.43 −0.91 −11.01 196.27 −2.567
(0.00) (0.06) (0.18) (0.00) (0.00) (0.00)
itcbi trend −0.77 48.27 −1.37 −9.52 180.40 −4.155
(0.22) (0.62) (0.09) (0.00) (0.00) (0.00)
no trend −2.86 68.99 −3.53 −10.91 230.86 −6.322
(0.00) (0.06) (0.00) (0.00) (0.00) (0.00)
Notes: The null hypothesis for all tests is that all variables are non-stationary. Results in parenthesis
are p-values. For IPS the Ztbar, for Fisher the Fisher statistic and for Pesaran the Ztbar statistic is
reported. In the IPS and Pesaran test, cross-sectional means are removed. All tests are performed under
the assumption of one lag. Test results for other lag structures reject a I(0) relation, but show a I(1)
relation for all variables.
Table 5
Cointegration test results.
Variables Kao ADF Pedroni PP Pedroni ADF
Group Panel Group Panel
(1) Baseline −2.86 −2.18 −2.30 −2.13 −1.73
(0.00) (0.01) (0.01) (0.02) (0.04)
(2) cbi −2.85 −2.85 −3.25 −2.51 −2.71
(0.00) (0.00) (0.00) (0.01) (0.03)
(3) itcbi −2.93 −3.50 −2.85 −3.28 −2.88
(0.00) (0.00) (0.00) (0.00) (0.00)
Notes: The null hypothesis for all tests is that the dependent and in-
dependent variables are not cointegrated. Kao ADF is the Augmented
Dickey–Fuller Kao residual test. Pedroni PP and ADF are the Phillips–
Peron and Augmented Dickey–Fuller Pedroni residual tests. P -values
are in parentheses.
194 L. Nöh / European Economic Review 119 (2019) 179–198
Appendix C. Subsamples
Table 6
Subsample for 1990–2010.
Variables 1990-2010 h
Long-run coefficients
Debt-to-GDP ratio 0.086 ∗∗∗ 1.49
(0.032) (0.22)
Inflation −0.001 0.06
(0.004) (0.81)
GDP growth rate −0.015 ∗∗∗ 6.60
(0.004) (0.01)
Long-term rate −0.042 ∗∗∗ 0.00
(0.005) (0.95)
Error correction term −0.428 ∗∗∗
(0.068)
Countries 28
Observations 476
Notes: Pooled mean group estimates for log of aver-
age maturity including controls for country and time
trend effects. Debt-to-GDP ratio is in log. Robust SE are
in parentheses for the coefficients. Significance is indi-
cated by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1. The Haus-
man test h rejects significant differences to the MG es-
timator. For the Hausman test, p -values are in paren-
theses. The dataset had to be expanded in a few places
with assumptions in order to obtain sufficiently long
observation periods for each country. These are the av-
erage maturities for Malta in 1998 and for Slovenia and
Switzerland in 1998 and 1999. For Malta and Slovenia,
the additional data are not fully compatible data from
ECB. For Switzerland, the two data points were chosen
as a precaution so that they correspond to a negative
trend between debt ratio and maturity.
Table 7
Subsamples for Euro and Non-Euro Countries, 1990–2016.
Variables Euro h Non-euro h
Long-run coefficients
Debt-to-GDP ratio 0.383 ∗∗∗ 0.92 0.076 1.26
(0.050) (0.34) (0.048) (0.26)
Inflation −0.033 ∗∗∗ 0.00 0.027 ∗∗∗ 1.23
(0.011) (0.98) (0.006) (0.27)
GDP growth rate −0.007 0.88 −0.038 ∗∗∗ 1.34
(0.005) (0.35) (0.007) (0.25)
Long-term rate −0.024 ∗∗∗ 0.51 −0.023 ∗∗∗ 1.16
(0.007) (0.48) (0.008) (0.28)
Error correction term −0.357 ∗∗∗ −0.486 ∗∗∗
(0.083) (0.094)
Countries 14 14
Observations 322 319
Notes: Pooled mean group estimates for log of average maturity including
controls for country and time trend effects. Debt-to-GDP ratio is in log.
Robust SE are in parentheses for the coefficients. Significance is indicated
by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1. The Hausman test h rejects signifi-
cant differences to the MG estimator. For the Hausman test, p -values are
in parentheses.
L. Nöh / European Economic Review 119 (2019) 179–198 195
Appendix D. Further robustness checks
Table 8
Subsamples for high and low CBI countries.
Variables CBI > 0.64 h CBI < 0.64 h
Long-run coefficients
Debt-to-GDP ratio 0.230 ∗∗∗ 0.88 0.067 0.02
(0.039) (0.35) (0.057) (0.88)
Inflation −0.019 ∗∗∗ 1.72 0.028 ∗∗∗ 0.00
(0.009) (0.19) (0.009) (0.96)
GDP growth rate −0.015 ∗∗∗ 3.02 −0.019 ∗∗∗ 0.01
(0.005) (0.08) (0.008) (0.93)
Long-term rate −0.039 ∗∗∗ 0.50 −0.009 0.77
(0.005) (0.48) (0.013) (0.38)
Error correction term −0.491 ∗∗∗ −0.364 ∗∗∗
(0.090) (0.075)
Countries 19 9
Observations 417 224
Notes: Pooled mean group estimates for log of average maturity including
controls for country and time trend effects. Debt-to-GDP ratio is in log.
Robust SE are in parentheses for the coefficients. Significance is indicated
by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1. The Hausman test h rejects signifi-
cant differences to the MG estimator. For the Hausman test, p -values are
in parentheses. 0.64 is the average central bank independence score in
the sample.
Table 9
Relation between debt-to-GDP ratio and average maturity with one fixed lag.
Variables Baseline h Incl. CBI h Incl. CBIIT h
Long-run coefficients
Debt-to-GDP ratio 0.159 ∗∗∗ 0.65 0.157 ∗∗∗ 1.23 0.036 1.06
(0.031) (0.42) (0.037) (0.27) (0.061) (0.30)
Inflation 0.003 0.86 0.003 1.05 0.016 ∗∗∗ 0.34
(0.010) (0.35) (0.010) (0.31) (0.008) (0.56)
GDP growth rate −0.055 ∗∗∗ 0.68 −0.045 ∗∗∗ 0.56 −0.037 ∗∗∗ 2.46
(0.008) (0.41) (0.008) (0.46) (0.006) (0.12)
Long-term rate −0.062 ∗∗∗ 1.38 −0.081 ∗∗∗ 0.25 −0.082 ∗∗∗ 0.22
(0.011) (0.24) (0.012) (0.62) (0.010) (0.64)
CBI 0.111 0.53 −0.956 ∗∗∗ 1.13
(0.099) (0.47) (0.378) (0.29)
CBIIT 0.245 ∗∗∗ 1.04
(0.081) (0.31)
Error correction term −0.289 ∗∗∗ −0.270 ∗∗∗ −0.292 ∗∗∗
(0.041) (0.043) (0.051)
Countries 28 28 28
Observations 641 641 641
Notes: Pooled mean group estimates for log of average maturity including controls for country
and time trend effects. Debt-to-GDP ratio is in log. CBI is a measure of central bank independence.
CBIIT is the interaction term between central bank independence and debt-to-GDP ratio.. Robust
SE are in parentheses for the coefficients. Significance is indicated by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗
p < 0.1. The Hausman test h rejects significant differences to the MG estimator. For the Hausman
test, p -values are in parentheses.
196 L. Nöh / European Economic Review 119 (2019) 179–198
Table 10
Additional interest rate controls.
Variables Short Rate h Spread h Spread and CBI h Spread and CBIIT h
Long-run coefficients
Debt-to-GDP ratio 0.147 ∗∗∗ 0.73 0.128 ∗∗∗ 0.64 0.129 ∗∗∗ 1.71 0.056 ∗∗∗ 0.52
(0.032) (0.39) (0.030) (0.43) (0.028) (0.19) (0.045) (0.47)
Inflation −0.003 0.87 −0.027 ∗∗∗ 0.00 −0.008 0.98 0.004 0.34
(0.005) (0.35) (0.006) (0.96) (0.005) (0.32) (0.003) (0.56)
GDP growth rate −0.007 ∗∗ 1.25 −0.012 ∗∗∗ 0.06 −0.011 ∗∗∗ 0.96 −0.009 ∗∗∗ 0.05
(0.004) (0.26) (0.004) (0.80) (0.004) (0.33) (0.003) (0.83)
Long-term rate −0.014 ∗∗ 0.91
(0.007) (0.34)
Interest rate spread 0.011 ∗∗ 0.48 0.013 ∗∗∗ 0.93 −0.010 ∗∗∗ 0.27
(0.006) (0.49) (0.005) (0.34) (0.003) (0.60)
Monetary policy rate −0.026 ∗∗∗ 0.88
(0.006) (0.35)
CBI 0.156 ∗∗∗ 1.42 -0.680 ∗∗∗ 0.19
(0.057) (0.23) (0.252) (0.67)
CBI-IT 0.211 ∗∗∗ 0.44
(0.059) (0.51)
Error correction term −0.423 ∗∗∗ −0.413 ∗∗∗ −0.494 ∗∗∗ −0.520 ∗∗∗
(0.065) (0.064) (0.070) (0.075)
Countries 28 28 28 28
Observations 641 641 641 641
Notes: Pooled mean group estimates for log of average maturity including controls country and time trend effects. Debt-to-GDP ratio
is in log. Interest rate spread is the difference between 10-year government bonds rate and the monetary policy rate. CBI is a measure
of central bank independence. CBIIT is the interaction term between central bank independence and debt-to-GDP ratio. Robust SE are
in parentheses for the coefficients. Significance is indicated by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1. The Hausman test rejects significant
differences to the MG estimator. The Hausman test h rejects significant differences to the MG estimator. For the Hausman test, p -values
are in parentheses.
Appendix E. Debt held by non-residents
Table 11
Share of debt held by non-residents.
Variables Foreign Debt h
Long-run coefficients
Debt-to-GDP ratio 1.87 ∗∗∗ 0.79
(0.027) (0.38)
Inflation -0.003 2.91
(0.004) (0.09)
GDP growth rate -0.012 ∗∗∗ 0.12
(0.004) (0.72)
Long-term rate −0.037 ∗∗∗ 3.07
(0.006) (0.08)
Non-resident debt 0.001 0.00
(0.001) (0.98)
Error correction term −0.500 ∗∗∗
(0.077)
Countries 28
Observations 607
Notes: Pooled mean group estimates for log of aver-
age maturity including controls for country and time
trend effects. Debt-to-GDP ratio is in log. Robust SE
are in parentheses for the coefficients. Significance
is indicated by: ∗∗∗ p < 0.01, ∗∗ p < 0.05, ∗ p < 0.1. The
Hausman test h rejects significant differences to the
MG estimator. For the Hausman test, p -values are in
parentheses. Non-resident debt is the share of debt
held by non-residents.
L. Nöh / European Economic Review 119 (2019) 179–198 197
020
4060
8010
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Australia
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Austria
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8010
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Belgium
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8010
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Canada0
2040
6080
100
1990
1995
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Czech Republic
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Denmark
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France
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Germany
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Greece
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Hungary
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Iceland
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Ireland
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Israel
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Malta
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Netherlands
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New Zealand
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Portugal
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Poland
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Slovakia
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Slovenia
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Spain
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Sweden
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Switzerland
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United Kingdom
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United States
Fig. 5. Share of Debt Held by Non-Residents, 1990–2016.
Note: Data from national authorities, OECD, and own calculations.
198 L. Nöh / European Economic Review 119 (2019) 179–198
Supplementary material
Supplementary material associated with this article can be found, in the online version, at doi: 10.1016/j.euroecorev.2019.
07.007 .
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