- Email: [email protected]
Assessing Ricardian equivalence for the New Member States: Does debt-neutrality matter?
Economic Systems 32 (2008) 119–128
Contents lists available at ScienceDirect
Economic Systems jo urnal homepage: www.elsevier.com/loc ate/ecos ys
Assessing Ricardian equivalence for the New Member States: Does debt-neutrality matter? Gerhard Reitschuler Department of Economics, University of Innsbruck, Universita¨tsstrasse 15, A-6020 Innsbruck, Austria
A R T I C L E I N F O
A B S T R A C T
Article history: Received 9 March 2007 Received in revised form 30 July 2007 Accepted 1 October 2007
Using a structural model based on dynamic optimizing agents, we empirically test the Ricardian equivalence proposition (REP) for 11 New EU-Member States (NMS). We extend the basic model by including the government budget constraint, thus being able to evaluate whether individuals take the evolution of public debt into account. In the basic setting we cannot reject the validity of the REP for four NMS, in the extended model the relevance of the REP changes for six countries, implying that the development of government debt and long-term sustainability of public finances matters with regard to the validity of the REP. ß 2008 Elsevier B.V. All rights reserved.
JEL classification: C32 E21 E62 Keywords: Ricardian equivalence Infinite horizons Liquidity constraints Government budget constraint Fiscal policy
1. Introduction The discussion concerning the size of the fiscal multiplier and the effect of fiscal policy on private consumption (and hence economic activity) has been discussed intensively during the past years. There are three main views with regard to this issue: traditional Keynesian theory predicts that – given a slack in productive capacity – there will be a rise in income (and hence private consumption) following an increase in government expenditure. Two theories contradict this doctrine: the Ricardian equivalence proposition (REP henceforward) states that an expansionary fiscal policy has no effect on consumption and output (i.e. the fiscal multiplier is equal to zero) since rational individuals – being aware of the intertemporal government budget constraint – will anticipate a future increase in taxes (necessary to
E-mail address: [email protected]. 0939-3625/$ – see front matter ß 2008 Elsevier B.V. All rights reserved. doi:10.1016/j.ecosys.2007.10.001
120
G. Reitschuler / Economic Systems 32 (2008) 119–128
balance the government budget constraint) by saving the respective amount. The Non-Keynesian theory even hypothesizes that, given a large budgetary imbalance, contractionary fiscal policy will induce a rise in consumption (i.e. the fiscal multiplier is negative) because it changes individuals’ expectations with regard to fiscal policy and hence about net lifetime income (Giavazzi and Pagano, 1990). In the context of the Maastricht Treaty, rising budget deficits and low GDP growth rates in the European Union (EU), the latter two theories have received more attraction lately. The reason is the following: every EU-country is subject to a deficit rule specified in the Maastricht Treaty which restricts public deficits to be lower than 3% of GDP.1 As GDP was growing at very moderate rates between 2001 and 2004 in the EU, some deficits came close to and over the 3%-threshold. Still, the European Commission repeatedly encouraged countries with ‘‘excessive’’ deficits to run contractionary fiscal policies (and at the same time suggested that the REP and the Non-Keynesian theory could hold)—exactly the opposite strategy suggested by the classical Keynesian theory. The only rationale to run such policies under these circumstances is if the REP and the Non-Keynesian theory were relevant—then expectations about the future play a role and in countries subject to the Maastricht rule expansionary fiscal policy would have no effect. Thus it is obvious to investigate whether the REP does indeed hold for EU-member-countries. As of May 2004 the EU has been enlarged by ten countries (the EU-10), as of January 2007 by two more, namely Romania and Bulgaria. Since these countries are also subject to the Maastricht fiscal rule this raises the issue whether, in the light of the general discussion outlined before, the REP is relevant. Another subject to be mentioned is the long-term sustainability of public finances: according to the European Commission (2005) several of the New Member States (NMS) will face substantially higher budgetary pressures due to population ageing in the upcoming decades, implying that the development of debt is relevant for the effectivity of fiscal policy. We are thus interested in two main questions: first, is there any relevance of the REP for the NMS and secondly, does the evolution of government debt play a role with regard to individuals’ consumption decisions? Following the seminal paper by Barro (1974), numerous studies have investigated the relevance of the REP; the empirical evidence is far from clear-cut. Tests of the REP have been performed by, e.g. estimating aggregate consumption functions (reduced-form consumption functions and Eulerequation type of specifications) or quantifying the effect of deficits on interest rates. Subject to these tests were either single countries or country aggregates, such as the U.S. (e.g. Feldstein, 1982; Seater and Mariano, 1985; Bernheim, 1987; Gale and Orszag, 2004), developing countries (Haque and Montiel, 1989; Khalid, 1996; Darius, 2001) or advanced economies (Masson et al., 1995; Giavazzi et al., 2000; Evans, 1993). For the NMS there is very little empirical literature testing for the REP.2 This piece of research contributes to the existing empirical literature in several aspects. We test for the relevance of the REP for the NMS, which in our opinion has not been sufficiently investigated so far3 using the structural model by Leiderman and Razin (1988) and Khalid (1996), modifying the basic model by including the government budget constraint. This allows us to not only to assess whether the REP is relevant but also to evaluate whether the evolution of government debt plays a role for individuals’ consumption decisions (i.e. whether government debt is net wealth). The paper is organized as follows: Section 2 introduces the basic as well as the modified model. Section 3 shows the results of the estimations of both models and then attempts to evaluate whether the development of debt and long-run sustainability of public finances play a role for the relevance of the REP. Section 4 concludes. 2. The theoretical model 2.1. The basic model In general, the REP is based on comparatively strong assumptions which may render its practical relevance, at least in its perfect form, questionable. The main premises of the REP are: perfect capital 1
Furthermore, public debt must not be higher than 60% of GDP unless it approaches the reference value ‘‘sufficiently fast’’. See, for example, Kaadu and Uusku¨la (2004). There is, however, more empirical evidence in context with twin deficits, see Fidrmuc (2003) and Aristovnik (2006). 3 One reason being the relatively short time series available for the NMS. 2
G. Reitschuler / Economic Systems 32 (2008) 119–128
121
markets (i.e. individuals are supposed to be able to lend and borrow freely, implying the absence of liquidity constraints), lump-sum taxes (i.e. taxes are non-distortionary) and very long-term planning horizons which in these type of models are mostly characterized as infinite lives (i.e. generations are linked with each other through bequests where the old generation cares for the young one). Each of these premises has its specific deficiencies with regard to its practical relevance, which may even more hold true for countries at a developing or transitional stage. Thus, for instance, capital markets are not fully developed in these countries, hence rendering the assumption of perfect capital markets debatable. Bearing in mind these restrictions there still is some empirical literature which suggests that the REP and its underlying assumptions may hold for countries in a transition stage (see, for instance, Khalid, 1996). The basic theoretical setting to the empirical study is the model proposed by Khalid (1996) and Leiderman and Razin (1988). The expected life-time utility of the liquidity-unconstrained representative consumer at time t is given by Et
1 X j u ðgdÞ Uðctþ jÞ
(1)
j¼0
where Et is the conditional expectations operator at time t, cu t denotes effective consumption of unconstrained individuals (i.e. individuals facing no liquidity constraints, denoted by superscript u), g is the probability of survival from one period to the next4 and d is a subjective discount factor. Following Bailey (1971) and Karras (1994), we assume that cu t is a combination of public consumption u gt and private consumption cut (cu t ¼ ct þ sgt , with s indicating the degree of substitutability between private and public consumption). The individual maximizes (1) subject to the following lifetime budget constraint, ctu ¼ but þ yut
R
g
but1 þ s g t
(2)
where bt is a real one period bond and yt is real income net of taxes (both of unconstrained individuals), R = 1 + r where r is the (risk-free) real interest rate (assumed constant) and (R/g) is the effective interest rate adjusted for life-time uncertainty. Making use of a no-Ponzi-game-rule ðlim ðg=RÞbut ¼ 0Þ, we substitute forward (2), which yields the t!1 intertemporal budget constraint: Et
1 j X g j¼0
R
ctu ¼ Et
1 j X g j¼0
R
R u yutþ j þ s gtþ but1 ¼ Et wut j
g
(3)
where Et wut denotes expected wealth. According to (3), the present discounted value of real consumption equals the present discounted value of real income net of taxes plus the weight attached to public consumption less the real interest paid on the last period’s debt commitment. Assuming a quadratic utility function, Uðct Þ ¼ act
1 2 c 2 t
it can be shown that the solution to the maximization problem takes the form:5
ctu ¼ b0 þ b1 Et wut
(4)
with:
b0 ¼
ag ð1 dRÞ dRðR g Þ
and
b1 ¼ 1
g dR2
4 For simplicity, this probability is assumed to be independent of age. Thus the probability that the individual survives at t t periods is g . 5 For a detailed derivation see Leiderman and Razin (1988).
122
G. Reitschuler / Economic Systems 32 (2008) 119–128
According to (4), individuals spend a proportion of wealth for consumption, i.e. they behave in accordance with the permanent-income hypothesis. We will now change an initial conjecture assuming that a fraction of the population is liquidityconstrained. The liquidity-constrained group receives a fixed proportion (u) of total labor income and entirely spends current income on consumption ðCtc ¼ Ytc Þ, with superscript c denoting constrained households, whereas the unconstrained group gets (1 + u) of total income but behaves in permanentincome fashion, i.e. as in Eq. (4). It can then be shown that aggregate consumption, Ct, defined as the sum of consumption of the liquidity-constrained and the liquidity-unconstrained group (C t ¼ Ctc þ Ctu ), is given by u þ ð1 g Þð1 u Þb1 Et1 Htu þ ð1 g Þb1 Et1 s St s Gt C t ¼ ð1 RÞb0 þ ð1 b1 ÞRCt1
þ uY t þ ð1 b1 ÞRðuY t s Gt1 Þ þ ðb1 g ½eyt þ egt Þ eyt
(5)
egt
and are the expectation errors for income and government consumption made by where individuals, Htu is human wealth and St non-human wealth, defined as Et1 Htu ¼ Et1
1 t X g
t ¼0
R
u Ytþ j
and Et1 St ¼ Et1
1 t X g
t¼0
R
Gutþ j
We will further assume that income and government spending follow an ARIMA (1, 1, 0) process, so that:
DY t ¼ r1 DY t1 þ hYt and DGt ¼ r1 DGt1 þ hGt where hYt and hG t are iid errors, independent from each other. Using the above processes and substituting values of b0 and b1 yields the following reduced-form equation: C t ¼ l0 þ l1 C t1 þ l2 Y t1 þ l3 Y t2 þ l4 Gt1 þ l5 Gt2 þ yt
(6)
where yt is the error term, assumed to be iid normal. Eq. (6) states that private consumption is determined by an ARDL (1, 2, 2) process without contemporaneous effects of government spending and income, l1 to l5 are the structural parameters, given by
l0 ¼
ag ð1 RÞð1 dRÞ dRðR g Þ
(7)
l1 ¼
g dR
(8)
"
g g þ ð1 uÞð1 g Þ 1 2 l2 ¼ u 1 þ r1 dR dR
R2 ð1 þ r1 Þ Rgr1 ðR g ÞðR gr1 Þ
# ! g R 2 r1 l3 ¼ ð1 uÞðg 1Þ 1 2 ur1 ðR g ÞðR gr1 Þ dR
!# (9)
"
l4 ¼
" g
!# 2 g R ð1 þ r2 Þ Rgr2 1 r2 þ ð1 g Þ 1 2 s ðR g ÞðR gr2 Þ dR dR
"
l5 ¼ r2 þ ðg 1Þ 1
g dR2
R2 ð1 þ r2 Þ Rgr2 ðR g ÞðR gr2 Þ
(10)
(11)
!#
s
(12)
G. Reitschuler / Economic Systems 32 (2008) 119–128
123
The constraints implied by the REP are those of infinite horizon (g = 1) and no liquidity constraints (u = 0). If these conditions are fulfilled, the lags of disposable income have no effect on current consumption. Furthermore, if the degree of substitutability between the private and public consumption spending is zero (s = 0), the model reduces to the Hall’s consumption model (Hall, 1978), implying that consumption follows a random walk. If, however, individuals behave as if they had finite lives (g < 1) and/or a strictly positive proportion of income is owned by liquidity-constrained agents (u > 0), fiscal policy measures will have an effect on expected income and private consumption. 2.2. The extended model In this section the aggregate consumption function given by (6) is augmented with the government budget constraint. This allows us to evaluate whether the evolution of public debt is relevant for the validity of the REP. The one-period government budget constraint in period t is given by:
t t ¼ g t bt þ Rbt1
(13)
where tt are government tax receipts, gt is government consumption, bt is real government debt (all in period t). Substituting forward Eq. (13) gives the intertemporal budget constraint:6 j 1 Bt j j!1 R
Et T t ¼ Et Gt þ RBt1 lim
(14)
with: Et T t ¼ Et
1 X 1 j¼0
R
T tþ j
and Et Gt ¼ Et
1 X 1 G R tþ j j¼0
Imposing a no-Ponzi-game rule ðlim ð1=RÞ j Bt j ¼ 0Þ yields: j!1
Et T t ¼ Et Gt þ RBt1
(15)
Eq. (15) implies that a deficit-financed cut in current taxes leads to higher future taxes. The existence of debt neutrality can be tested by substituting the government budget constraint into (5): u C t ¼ ð1 RÞb0 þ ð1 b1 ÞRCt1 þ ð1 g Þð1 u Þb1 Et1 Ht þ ð1 g Þb1 Et1 ðs 1ÞSt
s Gt þ uY t þ ð1 b1 ÞRðuY t s Gt1 Þ b1 ð1 g ÞRBt1 þ b1 g ½eyt þ egt
(16)
Rearranging and substituting for unobservables yields the following reduced form equation: C t ¼ l0 þ l1 C t1 þ l2 Y t1 þ l3 Y t2 þ l4 Gt1 þ l5 Gt2 þ l6 Bt1 þ yt
(17)
where l0 to l3 are equivalent to (7)–(10), and l4 to l6 are given by:
l4 ¼
" g
!# g s 1 R2 ð1 þ r2 Þ Rgr2 1 r2 þ ð1 g Þ 1 2 s s ðR g ÞðR gr2 Þ dR dR
!# g s 1 R2 ð1 þ r2 Þ Rgr2 l5 ¼ r2 þ ðg 1Þ 1 2 s s ðR g ÞðR gr2 Þ dR
(18)
"
6
Note that the discount rate is 1/R instead of g/R due to an infinite planning horizon of the government.
(19)
124
G. Reitschuler / Economic Systems 32 (2008) 119–128
g l6 ¼ R ð1 g Þ 1 2 dR
(20)
Again, the key parameters for the REP to hold are g and u: with g = 1 and/or u = 0, individuals behave Ricardian and hence consider today’s deficit financing as tomorrow’s tax liabilities. 3. Econometric methodology 3.1. Ricardian equivalence in the New Member States: the basic model For the empirical analysis we use yearly data on private consumption (Ct), government consumption (Gt), per-capita income (Yt)7 and the real interest rate (r) from 1990 to 2006 for eleven NMS.8 Government debt (Dt) is only available from 1995 onwards, which implies that we have to interpret the results for the extended model with caution. The data used, namely the time series for private consumption, government consumption, per-capita income, the real interest rate and government debt were all obtained from the European Commission’s Ameco database.9 As a first step we estimate the structural parameters implied by (7)–(12).10 Given the complex set of parameter restrictions, we will exogenously fix the values of some parameters of the model. The parameter corresponding to the risk-free real interest rate, R, was calculated as the average long-term real interest rate during the period 1990–2006. The parameters corresponding to the ARIMA representation of Yt and Gt (r1 and r2) are obtained by estimating the univariate models on Yt and Gt using all available observations. After imposing these parameters, the original model is left with five structural parameters to be estimated (a, g, d, u and s). The estimation of (7)–(12) was performed using full information maximum likelihood (see Table 1).11 The results render g highly significant and close to one for practically all countries. The estimates of u are significantly different from zero – implying that at least some fraction of the population is liquidityconstrained – for Estonia, Hungary, Malta and Poland. The parameter estimate for s is non-significant for all countries, suggesting that private and public consumption are no substitutes. The last column of Table 1 shows the likelihood ratio test statistic for the null hypothesis of g = 1 and u = 0, i.e. the premises implied by the REP of infinite horizons and no liquidity constraints. The likelihood ratio test indicates significant deviations from the REP for seven countries. For the remaining four countries, Cyprus, Latvia, the Slovak Republic and Slovenia, the null cannot be rejected at standard levels of significance.12 Given the comparative short sample period we have also estimated pooled regressions – based on different assumptions for the error term – for the countries of our sample. The first regression shows the result for the assumption that the error term is independent of the crosssectional units (countries) and iid normal (that is, the panel is estimated as if it were a crosscountry regression including a constant), the second one shows the results for the assumption of fixed country effects. Both regressions indicate significant deviations from the REP. We also estimated pooled regressions for country groups with similar characteristics (either per-capita GDP or country size). Group 1 comprises the Baltic countries (Estonia, Latvia and Lithuania), group 2 comprises the Czech Republic, Hungary and the Slovak Republic and group 3 Cyprus, Malta and Slovenia.13 The results confirm those obtained for single countries: the REP 7 Data is not (sufficiently) available for disposable income; hence, as suggested by Flavin (1985), we use per-capita income as a proxy. 8 We did not include Bulgaria due to incomplete time series on government debt. 9 Notice that, given the nonstationary nature of the variables included in the analysis, the ARDL(1,2,2) specification admits an error correction representation and a cointegration relationship between Ct, Yt and Gt is assumed to exist. Results concerning the stationarity of the deviations from the long-run equilibrium are available upon request. 10 Results from the unrestricted model implied by Eq. (6) (which does not allow a direct test of the REP) are available upon request. 11 Due to lack of comparability, the estimates of the utility function parameter a are not reported in the table, but are available upon request. 12 We have also estimated the baseline model using quarterly data. Using quarterly data does not change the results substantially. Results are available from the authors upon request. 13 Cyprus, Malta and Slovenia are pooled given their similar per capita GDP and despite the fact that Cyprus and Malta are significantly smaller in country size than Slovenia.
G. Reitschuler / Economic Systems 32 (2008) 119–128
125
Table 1 Parameter estimates for the restricted model
g
d
u
s
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
1.03 (0.72) 0.97*** (0.04) 1.00*** (0.00) 1.00*** (0.00) 1.00*** (0.30) 1.02** (0.54) 1.02*** (0.01) 1.04*** (0.01) 1.01*** (0.00) 1.01*** (0.04) 1.01*** (0.28)
0.96*** (0.00) 2.25 (1.78) 1.72 (1.42) 0.86*** (0.25) 0.99*** (0.28) 0.96*** (0.26) 0.97*** (0.35) 1.08*** (0.04) 1.49*** (0.25) 0.98*** (0.42) 0.98*** (0.05)
0.00 (1.75) 10.34 (1.78) 0.84*** (0.18) 1.28*** (0.18) 0.53 (13.96) 2.24* (1.17) 0.24** (0.12) 1.00*** (0.12) 0.64 (5.31) 1.16 (28.23) 0.08 (3.71)
1.13 0.60 0.96 2.53 0.80 0.49 0.62 0.00 0.05 0.60 1.81
EU-111 EU-112 Group 1 Group 2 Group 3
1.01*** 1.00*** 1.05*** 0.99*** 1.01***
0.95*** (0.05) 1.01*** (0.00) 0.95*** (0.05) 1.04** (0.52) 0.99*** (0.10)
0.91*** (0.01) 1.00*** (0.00) 1.11*** (0.35) 1.77 (2.62) 0.17 (0.16)
(0.00) (0.01) (0.00) (0.00) (0.02)
(4.50) (0.36) (2.15) (2.11) (6.91) (2.64) (0.89) (0.00) (0.74) (21.07) (7.91)
1.11*** (0.14) 1.56*** (0.30) 0.86 (2.35) 0.90 (7.04) 0.69 (0.70)
LRT
R2adj
DW
Obs
0.00 13.84*** 117.95*** 86.83*** 0.00 4.64*** 2.33E+06*** 2.62E+08*** 4.54E+04*** 3.94 0.11
0.94 0.95 0.98 0.98 0.97 0.98 0.96 0.98 0.95 0.98 0.98
2.19 2.15 2.00 2.21 0.88 2.54 1.43 2.52 2.21 2.77 2.07
15 15 14 14 14 14 15 15 15 14 15
1.04E+04*** 9.44E+04*** 17.14*** 0.47 1.65
0.94 0.94 0.96 0.89 0.92
1.33 1.91 1.47 1.73 1.86
161 161 39 39 45
The dependent variable is private consumption in all specifications. Time period: 1990–2006. White heteroskedasticity/serial correlation-corrected standard errors in parenthesis. ***(**)[*] stands for 1% (5%) [10%] significant, LRT is the likelihood-ratio test with the null of Ricardianity. EU-11 is the pooled regression for all countries of the sample with: (1) common intercept, (2) fixed effects; group 1 comprises EST, LVA, LTU, group 2 CZE, HUN, SVK, group 3 CYP, MLT, SVN; DW refers to the Durbin-Watson test statistic; Obs refers to the number of observations.
is rejected for the Baltic States whereas for the other two country groups it cannot be rejected. 3.2. Does debt-neutrality matter? In general, the assumption that consumers internalize the government budget constraint ‘‘imposes formidable requirements on agents’ ability to gather and process information’’ (Lopez et al., 2000). However, the hypothesis that private agents take the government budget constraint into account with regard to their consumption decisions might still be a good approximation of reality. In this respect, the level as well as the increase of government debt may play a decisive role. It is thus probable that individuals develop awareness for the government budget constraint in countries where public debt follows a rapidly rising (or even explosive) path. Then consumers may become aware of future tax implications since a future fiscal adjustment is more likely (Pozzi, 2003, 2006).14 Generally, the evolution of the debt-to-GDP ratio in the NMS shows a somewhat heterogeneous picture. Table 2 gives the development of public debt since 1995 in 3–4-year averages. As can be seen there are mainly three groups of countries: one group (Estonia, Latvia and Slovenia) displays a relatively stable debt evolution, another group (Lithuania and the Slovak Republic) has succeeded in cutting public debt, sometimes even drastically. Yet another group (Cyprus, Malta and Poland) has experienced the opposite, i.e. rapidly rising debt ratios. A second issue to be mentioned is the long-run sustainability of public finances:15 it should be noted that demographic projections are particularly worrisome in the NMS.16 According to the 14
This notion is mentioned, for instance, by Nicoletti (1988, 1992) and Dalamagas (1993). There are different definitions with regard to long-term sustainability of public finances. We use the one of the European Commission (see European Commission, 2006) that defines a sustainable public finance position as to ‘‘whether on the basis of current policies and projected budgetary trends Member States will: (i) meet the government’s intertemporal budget constraint so that the discounted value of future revenues matches the discounted value of future government expenditures and the level of outstanding debt; and, (ii) continue to comply with the budgetary requirements of EMU, and in particular, the Treaty requirement to keep debt levels below the 60 percent of GDP reference value.’’ Regarding the issue of sustainability in transition countries see also Aristovnik and Bercˇicˇ (2007). 16 This is due to the fact that most countries have relatively low fertility rates, a lower life expectancy than EU-15, and a negative migration balance (see European Commission, 2005). 15
G. Reitschuler / Economic Systems 32 (2008) 119–128
126
Table 2 Evolution of Government Debt-to-GDP ratio in the New Member States Country
AVG 1996–1998
AVG 1999–2002
AVG 2003–2006
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
28.1 13.4 6.5 49.4 13.3 51.1 46.1 26.9 17.1 59.1 28.9
61.7 22.3 5.4 55.3 13.1 23.0 58.8 37.7 23.7 47.4 27.6
70.2 30.7 5.0 58.0 13.1 19.6 74.1 43.5 17.2 38.3 29.4
AVG refers to the average value of public debt for the period indicated; data source is the Ameco Database (European Commission).
European Commission (2005), long-run budgetary pressures due to population ageing (as expressed by the average old-age dependency ratio) are projected to more than double in Cyprus, the Czech Republic, Poland and Slovenia and even triple in the Slovak Republic by 2050. These two issues – the development of the debt ratio over the last decade and long-run sustainability of public finances – make the investigation of a model with government debt worthwhile: if, for instance, the evolution of debt becomes unsustainable, then an expansionary policy will have no effect on output (i.e. the fiscal multiplier is zero). On the other hand, countries curbing government debt should be able to reap the laurels of such sustainable fiscal policies sooner or later. The aim of this section is thus to evaluate whether individuals take these developments into account in their consumption decisions. Table 3 shows the following results for the extended model: the likelihood ratio test for the null hypothesis of g = 1 and u = 0 indicates significant deviations from the REP for five countries. For the remaining six countries (Cyprus, the Czech Republic, Malta, Poland, Romania and the Slovak Republic) the null cannot be rejected at standard levels of significance. Again, pooled regressions were estimated due to the fact that the time series for the countries in question which are presented in the last rows of Table 3 are comparatively short. Both regressions for the whole sample indicate significant deviations Table 3 Parameter estimates for the modified restricted model
g
d
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
1.03** (0.49) 1.92 (1.43) 1.00*** (0.00) 0.98*** (0.00) 1.03*** (0.24) 1.04*** (0.05) 1.00*** (0.01) 1.10*** (0.28) 0.99*** (0.01) 1.01** (0.49) 1.01*** (0.00)
EU-111 EU-112 Group 1 Group 2 Group 3
1.07*** 1.06*** 0.99*** 0.99*** 1.00***
(0.02) (0.00) (0.00) (0.00) (0.00)
0.96*** (0.10) 1.75 (1.49) 2.60 (13.63) 0.97*** (0.27) 0.92** (0.42) 0.96 (0.63) 1.13*** (0.10) 1.17** (0.57) 1.52*** (0.58) 0.98*** (0.37) 0.98*** (0.00) 1.08*** (0.03) 1.04*** (0.00) 0.77 (2.49) 0.03 (0.57) 0.88*** (0.04)
u
s
LRT
R2adj
DW
Obs
0.48 (2.62) 0.28 (98.71) 0.78*** (0.42) 0.13 (246.20) 0.99 (2.96) 1.14 (0.78) 0.06 (0.43) 0.08 (6.50) 0.29 (12.27) 0.92 (20.22) 0.62 (1.13)
0.46 (3.73) 2.57 (6.36) 0.79 (2.62) 0.23 (0.77) 7.48 (15.55) 1.06 (1.14) 1.20 (1.21) 4.11 (4.99) 2.96* (1.77) 0.27 (4.49) 1.52 (1.21)
0.18 5.11* 52.39*** 13.39*** 6898.71*** 10.64*** 0.40 0.58 1.06 0.00 6.15**
0.91 0.92 0.98 0.98 0.98 0.98 0.91 0.98 0.96 0.98 0.98
2.20 2.17 1.64 2.27 1.98 2.59 2.13 2.51 2.07 2.57 1.82
11 11 11 11 11 11 11 11 10 11 11
0.70 (0.45) 0.39 (0.25) 0.18 (2.00) 0.20 (0.98) 0.13** (0.06)
2.62*** (0.42) 2.57*** (0.79) 0.78 (22.18) 0.01 (0.56) 1.01 (0.70)
11.73*** 405.26*** 0.04 0.47 45.49***
0.84 0.91 0.97 0.84 0.90
1.41 2.20 1.47 1.25 2.49
118 118 33 33 31
The dependent variable is private consumption in all specifications. Time period: 1995–2006. White heteroskedasticity/serial correlation-corrected standard errors in parenthesis. ***(**)[*] stands for 1% (5%) [10%] significant, LRT is the likelihood-ratio test with the null of Ricardianity. EU-11 is the pooled regression for all countries of the sample with: (1) common intercept, (2) fixed effects; group 1 comprises EST, LVA, LTU, group 2 CZE, HUN, SVK, group 3 CYP, MLT, SVN; DW refers to the Durbin-Watson test statistic; Obs refers to the number of observations.
G. Reitschuler / Economic Systems 32 (2008) 119–128
127
Table 4 Summary of the LR-test results for the model with and without government budget constraint Country
REP model I
REP model II
Debt Evolution
LTS
FS (st)
FS (lt)
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
0.00 13.84*** 117.95*** 86.83*** 0.00 4.64** 2.3E+06*** 2.6E+08*** 4.5E+04*** 3.94 0.11
0.18 5.11* 52.39*** 13.39*** 6898.71*** 10.64*** 0.40 0.58 1.06 0.00 6.15**
(Rapidly) rising Rising Constant Rising Constant (Rapidly) falling (Rapidly) rising (Rapidly) rising Rising-falling (Rapidly) falling Constant
No No (Yes) (Yes) (Yes) (Yes) (Yes) (No) (Yes) No No
n.a. No Yes No Yes Yes n.a. No Yes Yes Yes
n.a. No Yes No Yes Yes n.a. No Yes No No
The numbers refer to the results of the LR-Test, LTS (long-term sustainability) refers to the average dependency ratio as an indicator for fiscal pressures in the future (source: European Commission, 2005); FS(st) [FS(lt)] refers to the evaluation of shortterm [long-term] fiscal sustainability of Aristovnik and Bercˇicˇ (2007); ***(**)[*] stands for 1% (5%) [10%] significant.
from the REP. As in the section before we have estimated pooled regressions for country groups: we cannot reject the REP for the Baltic countries (Estonia, Latvia, Lithuania) and the group with the Czech Republic, Hungary and the Slovak Republic, whereas the REP is rejected for the country group comprising Cyprus, Malta and Slovenia. We now turn to the comparison of the results from the baseline and the extended model (see Table 4). It shows a change in the validity of the REP from Non-Ricardian (in the baseline model) to Ricardian (in the extended model) for the Czech Republic, Malta, Poland, Romania and a change from Ricardian to Non-Ricardian for Latvia and Slovenia. These results are in accordance with the development of debt, long-run budgetary pressures and the evaluation of (short- and long-term) sustainability by Aristovnik and Bercˇicˇ (2007): the Czech Republic, Malta and Poland (which are all Ricardian in the extended model) display rapidly rising debt ratios; Latvia and Slovenia (which are Non-Ricardian in the extended model) show a constant debt ratio. Obviously in the Slovak Republic (which still displays Ricardian behaviour in the extended model) long-run sustainability plays a more prominent role than the rapidly falling debt ratio in the past decade. Interestingly, all five countries that according to the European Commission (2005) will experience substantial age-related expenditures in the future (which would imply rising debt) are Ricardian in the extended model. 4. Conclusions This piece of research presents evidence concerning the validity of the REP for 11 new EU member states. We have investigated two main questions: first, whether the REP has any relevance for these NMS and secondly, if the evolution of debt plays a role with regard to individuals’ consumption decisions. We have investigated these two issues using the model by Leiderman and Razin (1988) and Khalid (1996), extending the basic model by including the government budget constraint. The baseline setting indicates significant deviations from the REP for 7 out of 11 countries (i.e. for all except Cyprus, Latvia, the Slovak Republic and Slovenia). In the extended model we cannot reject the validity of the REP for Cyprus, the Czech Republic, Malta, Poland, Romania and the Slovak Republic. A comparison of the results from the baseline and the extended model shows a change in the validity of the REP (either from Non-Ricardian to Ricardian or vice versa) for the Czech Republic, Latvia, Malta, Poland, Romania and Slovenia. These results are in line with the evolution of government debt and long-run sustainability in those countries: the Czech Republic, Malta and Poland (rapidly rising debt evolution) display a change from Non-Ricardian to Ricardian behaviour and Latvia and Slovenia (having a constant debt evolution) are Non-Ricardian after having been Ricardian in the baseline model. One policy implication for the NMS concerns the situation after EMU entry: one of the main prerequisites for EMU entry is the 3%-deficit criterion which will restrict the room for fiscal policy manoeuvre for some NMS after joining EMU. For other NMS, which already now display a deficit close
128
G. Reitschuler / Economic Systems 32 (2008) 119–128
to balance or even in surplus (such as the Baltic countries), there may be more room for expansive fiscal policy which, however, could be ineffective (with regard to consumption and thus growth) given that the REP holds. Hence, for those countries for which the REP holds, letting their automatic stabilizers play over the cycle and raising growth (potential) via structural reforms may be more appropriate as compared with implementing discretionary fiscal policy measures. Our results suggest that the NMS should closely watch the development of public debt and longrun sustainability of public finances. Clearly, when debt follows a rising or even explosive path the effectivity of fiscal measures will be reduced since individuals will expect future tax implications, thus rendering fiscal policy ineffective. Hence, reducing long-run budgetary pressures may be an instrument to raise the effectiveness of fiscal policy in the respective countries. References Aristovnik, A., 2006. The influence of fiscal policy and private investment on external imbalances in transition economies. Econ. Bus. Rev. 8, 143–166. Aristovnik, A., Bercˇicˇ, B., 2007. Fiscal sustainability in selected transition countries. MPRA Working Paper 122, Munich. Barro, R.J., 1974. Are government bonds net wealth? J. Polit. Econ. 82, 1095–1117. Bailey, M.J., 1971. National Income and the Price Level. McGraw-Hill, New York. Bernheim, D., 1987. Ricardian equivalence: an evaluation of the theory and evidence. In: Fisher, S. (Ed.), NBER Macroeconomics Annual 1987. MIT Press, Cambridge, MA, pp. 263–304. Dalamagas, B., 1993. How efficient is the substitution of debt for taxes in influencing demand? Appl. Econ. 25, 295–303. Darius, R., 2001. Debt neutrality: theory and evidence from developing countries. Appl. Econ. 33, 49–58. European Commission, 2005. Public finances in EMU 2005. European Economy 3, Brussels. European Commission, 2006. Public finances in EMU 2006. European Economy 3, Brussels. Evans, P., 1993. Consumers are not Ricardian: evidence from 19 countries. Econ. Inquiry 31, 534–548. Feldstein, M., 1982. Government deficits and aggregate demand. J. Monet. Econ. 9, 1–20. Fidrmuc, J., 2003. The Feldstein-Horioka puzzle and twin deficits in selected economies. Econ. Plan. 36, 135–152. Flavin, M., 1985. Excess sensitivity of consumption to current income: liquidity constraints or myopia? Can. J. Econ. 18, 117– 136. Gale, W., Orszag, P., 2004. Budget deficits, national saving, and interest rates. Brookings Papers Econ. Activ. 2, 101–210. Giavazzi, F., Pagano, M., 1990. Can severe fiscal contractions be expansionary? Tales of two small European countries. In: Blanchard, O., Fisher, S. (Eds.), NBER Macroeconomics Annual 1990. MIT Press, Cambridge, MA, pp. 75–111. Giavazzi, F., Jappelli, T., Pagano, M., 2000. Searching for non-linear effects of fiscal policy: evidence from industrial and developing countries. Eur. Econ. Rev. 44, 1259–1289. Hall, R.E., 1978. Stochastic implications of the life cycle-permanent income hypothesis: theory and evidence. J. Polit. Econ. 86, 971–987. Haque, N., Montiel, P., 1989. Consumption in developing countries: tests for liquidity constraints and finite horizons. Rev. Econ. Stat. 7, 408–415. Kaadu, H., Uusku¨la, L., 2004. Liquidity constraints and Ricardian equivalence in Estonia. Bank of Estonia Working Paper 2004, 7, Tallinn. Karras, G., 1994. Government spending and private consumption: some international evidence. J. Money Credit Bank. 26, 9–22. Khalid, A.M., 1996. Ricardian equivalence: empirical evidence from developing economies. J. Develop. Econ. 51, 413–432. Leiderman, L., Razin, A., 1988. Testing Ricardian neutrality with an intertemporal stochastic model. J. Money Credit Bank. 20, 1– 21. Lopez, H.J., Schmidt-Hebbel, K., Serve´n, L., 2000. How effective is fiscal policy in raising national saving? Rev. Econ. Stat. 82, 226– 238. Masson, P., Bayoumi, T., Samiei, H., 1995. Saving behaviour in industrial and developing countries. In: IMF (Eds.), Staff studies for the World economic outlook, World economic and financial surveys. IMF, Washington, DC, pp. 1–27. Nicoletti, G., 1988. Private consumption, inflation and the debt neutrality hypothesis: the case of eight OECD countries. OECD Working Paper 50, Paris. Nicoletti, G., 1992. Is tax-discounting stable over time? Oxford Bull. Econ. Stat. 54, 121–144. Pozzi, L., 2003. Tax discounting in a high-debt economy. Oxford Bull. Econ. Stat. 65, 261–282. Pozzi, L., 2006. Ricardian equivalence under imperfect information. J. Public Econ. 90, 2009–2026. Seater, J.J., Mariano, R.S., 1985. New tests of the life cycle and tax discounting hypotheses. J. Monetary Econ. 15, 195–215.
Contents lists available at ScienceDirect
Economic Systems jo urnal homepage: www.elsevier.com/loc ate/ecos ys
Assessing Ricardian equivalence for the New Member States: Does debt-neutrality matter? Gerhard Reitschuler Department of Economics, University of Innsbruck, Universita¨tsstrasse 15, A-6020 Innsbruck, Austria
A R T I C L E I N F O
A B S T R A C T
Article history: Received 9 March 2007 Received in revised form 30 July 2007 Accepted 1 October 2007
Using a structural model based on dynamic optimizing agents, we empirically test the Ricardian equivalence proposition (REP) for 11 New EU-Member States (NMS). We extend the basic model by including the government budget constraint, thus being able to evaluate whether individuals take the evolution of public debt into account. In the basic setting we cannot reject the validity of the REP for four NMS, in the extended model the relevance of the REP changes for six countries, implying that the development of government debt and long-term sustainability of public finances matters with regard to the validity of the REP. ß 2008 Elsevier B.V. All rights reserved.
JEL classification: C32 E21 E62 Keywords: Ricardian equivalence Infinite horizons Liquidity constraints Government budget constraint Fiscal policy
1. Introduction The discussion concerning the size of the fiscal multiplier and the effect of fiscal policy on private consumption (and hence economic activity) has been discussed intensively during the past years. There are three main views with regard to this issue: traditional Keynesian theory predicts that – given a slack in productive capacity – there will be a rise in income (and hence private consumption) following an increase in government expenditure. Two theories contradict this doctrine: the Ricardian equivalence proposition (REP henceforward) states that an expansionary fiscal policy has no effect on consumption and output (i.e. the fiscal multiplier is equal to zero) since rational individuals – being aware of the intertemporal government budget constraint – will anticipate a future increase in taxes (necessary to
E-mail address: [email protected]. 0939-3625/$ – see front matter ß 2008 Elsevier B.V. All rights reserved. doi:10.1016/j.ecosys.2007.10.001
120
G. Reitschuler / Economic Systems 32 (2008) 119–128
balance the government budget constraint) by saving the respective amount. The Non-Keynesian theory even hypothesizes that, given a large budgetary imbalance, contractionary fiscal policy will induce a rise in consumption (i.e. the fiscal multiplier is negative) because it changes individuals’ expectations with regard to fiscal policy and hence about net lifetime income (Giavazzi and Pagano, 1990). In the context of the Maastricht Treaty, rising budget deficits and low GDP growth rates in the European Union (EU), the latter two theories have received more attraction lately. The reason is the following: every EU-country is subject to a deficit rule specified in the Maastricht Treaty which restricts public deficits to be lower than 3% of GDP.1 As GDP was growing at very moderate rates between 2001 and 2004 in the EU, some deficits came close to and over the 3%-threshold. Still, the European Commission repeatedly encouraged countries with ‘‘excessive’’ deficits to run contractionary fiscal policies (and at the same time suggested that the REP and the Non-Keynesian theory could hold)—exactly the opposite strategy suggested by the classical Keynesian theory. The only rationale to run such policies under these circumstances is if the REP and the Non-Keynesian theory were relevant—then expectations about the future play a role and in countries subject to the Maastricht rule expansionary fiscal policy would have no effect. Thus it is obvious to investigate whether the REP does indeed hold for EU-member-countries. As of May 2004 the EU has been enlarged by ten countries (the EU-10), as of January 2007 by two more, namely Romania and Bulgaria. Since these countries are also subject to the Maastricht fiscal rule this raises the issue whether, in the light of the general discussion outlined before, the REP is relevant. Another subject to be mentioned is the long-term sustainability of public finances: according to the European Commission (2005) several of the New Member States (NMS) will face substantially higher budgetary pressures due to population ageing in the upcoming decades, implying that the development of debt is relevant for the effectivity of fiscal policy. We are thus interested in two main questions: first, is there any relevance of the REP for the NMS and secondly, does the evolution of government debt play a role with regard to individuals’ consumption decisions? Following the seminal paper by Barro (1974), numerous studies have investigated the relevance of the REP; the empirical evidence is far from clear-cut. Tests of the REP have been performed by, e.g. estimating aggregate consumption functions (reduced-form consumption functions and Eulerequation type of specifications) or quantifying the effect of deficits on interest rates. Subject to these tests were either single countries or country aggregates, such as the U.S. (e.g. Feldstein, 1982; Seater and Mariano, 1985; Bernheim, 1987; Gale and Orszag, 2004), developing countries (Haque and Montiel, 1989; Khalid, 1996; Darius, 2001) or advanced economies (Masson et al., 1995; Giavazzi et al., 2000; Evans, 1993). For the NMS there is very little empirical literature testing for the REP.2 This piece of research contributes to the existing empirical literature in several aspects. We test for the relevance of the REP for the NMS, which in our opinion has not been sufficiently investigated so far3 using the structural model by Leiderman and Razin (1988) and Khalid (1996), modifying the basic model by including the government budget constraint. This allows us to not only to assess whether the REP is relevant but also to evaluate whether the evolution of government debt plays a role for individuals’ consumption decisions (i.e. whether government debt is net wealth). The paper is organized as follows: Section 2 introduces the basic as well as the modified model. Section 3 shows the results of the estimations of both models and then attempts to evaluate whether the development of debt and long-run sustainability of public finances play a role for the relevance of the REP. Section 4 concludes. 2. The theoretical model 2.1. The basic model In general, the REP is based on comparatively strong assumptions which may render its practical relevance, at least in its perfect form, questionable. The main premises of the REP are: perfect capital 1
Furthermore, public debt must not be higher than 60% of GDP unless it approaches the reference value ‘‘sufficiently fast’’. See, for example, Kaadu and Uusku¨la (2004). There is, however, more empirical evidence in context with twin deficits, see Fidrmuc (2003) and Aristovnik (2006). 3 One reason being the relatively short time series available for the NMS. 2
G. Reitschuler / Economic Systems 32 (2008) 119–128
121
markets (i.e. individuals are supposed to be able to lend and borrow freely, implying the absence of liquidity constraints), lump-sum taxes (i.e. taxes are non-distortionary) and very long-term planning horizons which in these type of models are mostly characterized as infinite lives (i.e. generations are linked with each other through bequests where the old generation cares for the young one). Each of these premises has its specific deficiencies with regard to its practical relevance, which may even more hold true for countries at a developing or transitional stage. Thus, for instance, capital markets are not fully developed in these countries, hence rendering the assumption of perfect capital markets debatable. Bearing in mind these restrictions there still is some empirical literature which suggests that the REP and its underlying assumptions may hold for countries in a transition stage (see, for instance, Khalid, 1996). The basic theoretical setting to the empirical study is the model proposed by Khalid (1996) and Leiderman and Razin (1988). The expected life-time utility of the liquidity-unconstrained representative consumer at time t is given by Et
1 X j u ðgdÞ Uðctþ jÞ
(1)
j¼0
where Et is the conditional expectations operator at time t, cu t denotes effective consumption of unconstrained individuals (i.e. individuals facing no liquidity constraints, denoted by superscript u), g is the probability of survival from one period to the next4 and d is a subjective discount factor. Following Bailey (1971) and Karras (1994), we assume that cu t is a combination of public consumption u gt and private consumption cut (cu t ¼ ct þ sgt , with s indicating the degree of substitutability between private and public consumption). The individual maximizes (1) subject to the following lifetime budget constraint, ctu ¼ but þ yut
R
g
but1 þ s g t
(2)
where bt is a real one period bond and yt is real income net of taxes (both of unconstrained individuals), R = 1 + r where r is the (risk-free) real interest rate (assumed constant) and (R/g) is the effective interest rate adjusted for life-time uncertainty. Making use of a no-Ponzi-game-rule ðlim ðg=RÞbut ¼ 0Þ, we substitute forward (2), which yields the t!1 intertemporal budget constraint: Et
1 j X g j¼0
R
ctu ¼ Et
1 j X g j¼0
R
R u yutþ j þ s gtþ but1 ¼ Et wut j
g
(3)
where Et wut denotes expected wealth. According to (3), the present discounted value of real consumption equals the present discounted value of real income net of taxes plus the weight attached to public consumption less the real interest paid on the last period’s debt commitment. Assuming a quadratic utility function, Uðct Þ ¼ act
1 2 c 2 t
it can be shown that the solution to the maximization problem takes the form:5
ctu ¼ b0 þ b1 Et wut
(4)
with:
b0 ¼
ag ð1 dRÞ dRðR g Þ
and
b1 ¼ 1
g dR2
4 For simplicity, this probability is assumed to be independent of age. Thus the probability that the individual survives at t t periods is g . 5 For a detailed derivation see Leiderman and Razin (1988).
122
G. Reitschuler / Economic Systems 32 (2008) 119–128
According to (4), individuals spend a proportion of wealth for consumption, i.e. they behave in accordance with the permanent-income hypothesis. We will now change an initial conjecture assuming that a fraction of the population is liquidityconstrained. The liquidity-constrained group receives a fixed proportion (u) of total labor income and entirely spends current income on consumption ðCtc ¼ Ytc Þ, with superscript c denoting constrained households, whereas the unconstrained group gets (1 + u) of total income but behaves in permanentincome fashion, i.e. as in Eq. (4). It can then be shown that aggregate consumption, Ct, defined as the sum of consumption of the liquidity-constrained and the liquidity-unconstrained group (C t ¼ Ctc þ Ctu ), is given by u þ ð1 g Þð1 u Þb1 Et1 Htu þ ð1 g Þb1 Et1 s St s Gt C t ¼ ð1 RÞb0 þ ð1 b1 ÞRCt1
þ uY t þ ð1 b1 ÞRðuY t s Gt1 Þ þ ðb1 g ½eyt þ egt Þ eyt
(5)
egt
and are the expectation errors for income and government consumption made by where individuals, Htu is human wealth and St non-human wealth, defined as Et1 Htu ¼ Et1
1 t X g
t ¼0
R
u Ytþ j
and Et1 St ¼ Et1
1 t X g
t¼0
R
Gutþ j
We will further assume that income and government spending follow an ARIMA (1, 1, 0) process, so that:
DY t ¼ r1 DY t1 þ hYt and DGt ¼ r1 DGt1 þ hGt where hYt and hG t are iid errors, independent from each other. Using the above processes and substituting values of b0 and b1 yields the following reduced-form equation: C t ¼ l0 þ l1 C t1 þ l2 Y t1 þ l3 Y t2 þ l4 Gt1 þ l5 Gt2 þ yt
(6)
where yt is the error term, assumed to be iid normal. Eq. (6) states that private consumption is determined by an ARDL (1, 2, 2) process without contemporaneous effects of government spending and income, l1 to l5 are the structural parameters, given by
l0 ¼
ag ð1 RÞð1 dRÞ dRðR g Þ
(7)
l1 ¼
g dR
(8)
"
g g þ ð1 uÞð1 g Þ 1 2 l2 ¼ u 1 þ r1 dR dR
R2 ð1 þ r1 Þ Rgr1 ðR g ÞðR gr1 Þ
# ! g R 2 r1 l3 ¼ ð1 uÞðg 1Þ 1 2 ur1 ðR g ÞðR gr1 Þ dR
!# (9)
"
l4 ¼
" g
!# 2 g R ð1 þ r2 Þ Rgr2 1 r2 þ ð1 g Þ 1 2 s ðR g ÞðR gr2 Þ dR dR
"
l5 ¼ r2 þ ðg 1Þ 1
g dR2
R2 ð1 þ r2 Þ Rgr2 ðR g ÞðR gr2 Þ
(10)
(11)
!#
s
(12)
G. Reitschuler / Economic Systems 32 (2008) 119–128
123
The constraints implied by the REP are those of infinite horizon (g = 1) and no liquidity constraints (u = 0). If these conditions are fulfilled, the lags of disposable income have no effect on current consumption. Furthermore, if the degree of substitutability between the private and public consumption spending is zero (s = 0), the model reduces to the Hall’s consumption model (Hall, 1978), implying that consumption follows a random walk. If, however, individuals behave as if they had finite lives (g < 1) and/or a strictly positive proportion of income is owned by liquidity-constrained agents (u > 0), fiscal policy measures will have an effect on expected income and private consumption. 2.2. The extended model In this section the aggregate consumption function given by (6) is augmented with the government budget constraint. This allows us to evaluate whether the evolution of public debt is relevant for the validity of the REP. The one-period government budget constraint in period t is given by:
t t ¼ g t bt þ Rbt1
(13)
where tt are government tax receipts, gt is government consumption, bt is real government debt (all in period t). Substituting forward Eq. (13) gives the intertemporal budget constraint:6 j 1 Bt j j!1 R
Et T t ¼ Et Gt þ RBt1 lim
(14)
with: Et T t ¼ Et
1 X 1 j¼0
R
T tþ j
and Et Gt ¼ Et
1 X 1 G R tþ j j¼0
Imposing a no-Ponzi-game rule ðlim ð1=RÞ j Bt j ¼ 0Þ yields: j!1
Et T t ¼ Et Gt þ RBt1
(15)
Eq. (15) implies that a deficit-financed cut in current taxes leads to higher future taxes. The existence of debt neutrality can be tested by substituting the government budget constraint into (5): u C t ¼ ð1 RÞb0 þ ð1 b1 ÞRCt1 þ ð1 g Þð1 u Þb1 Et1 Ht þ ð1 g Þb1 Et1 ðs 1ÞSt
s Gt þ uY t þ ð1 b1 ÞRðuY t s Gt1 Þ b1 ð1 g ÞRBt1 þ b1 g ½eyt þ egt
(16)
Rearranging and substituting for unobservables yields the following reduced form equation: C t ¼ l0 þ l1 C t1 þ l2 Y t1 þ l3 Y t2 þ l4 Gt1 þ l5 Gt2 þ l6 Bt1 þ yt
(17)
where l0 to l3 are equivalent to (7)–(10), and l4 to l6 are given by:
l4 ¼
" g
!# g s 1 R2 ð1 þ r2 Þ Rgr2 1 r2 þ ð1 g Þ 1 2 s s ðR g ÞðR gr2 Þ dR dR
!# g s 1 R2 ð1 þ r2 Þ Rgr2 l5 ¼ r2 þ ðg 1Þ 1 2 s s ðR g ÞðR gr2 Þ dR
(18)
"
6
Note that the discount rate is 1/R instead of g/R due to an infinite planning horizon of the government.
(19)
124
G. Reitschuler / Economic Systems 32 (2008) 119–128
g l6 ¼ R ð1 g Þ 1 2 dR
(20)
Again, the key parameters for the REP to hold are g and u: with g = 1 and/or u = 0, individuals behave Ricardian and hence consider today’s deficit financing as tomorrow’s tax liabilities. 3. Econometric methodology 3.1. Ricardian equivalence in the New Member States: the basic model For the empirical analysis we use yearly data on private consumption (Ct), government consumption (Gt), per-capita income (Yt)7 and the real interest rate (r) from 1990 to 2006 for eleven NMS.8 Government debt (Dt) is only available from 1995 onwards, which implies that we have to interpret the results for the extended model with caution. The data used, namely the time series for private consumption, government consumption, per-capita income, the real interest rate and government debt were all obtained from the European Commission’s Ameco database.9 As a first step we estimate the structural parameters implied by (7)–(12).10 Given the complex set of parameter restrictions, we will exogenously fix the values of some parameters of the model. The parameter corresponding to the risk-free real interest rate, R, was calculated as the average long-term real interest rate during the period 1990–2006. The parameters corresponding to the ARIMA representation of Yt and Gt (r1 and r2) are obtained by estimating the univariate models on Yt and Gt using all available observations. After imposing these parameters, the original model is left with five structural parameters to be estimated (a, g, d, u and s). The estimation of (7)–(12) was performed using full information maximum likelihood (see Table 1).11 The results render g highly significant and close to one for practically all countries. The estimates of u are significantly different from zero – implying that at least some fraction of the population is liquidityconstrained – for Estonia, Hungary, Malta and Poland. The parameter estimate for s is non-significant for all countries, suggesting that private and public consumption are no substitutes. The last column of Table 1 shows the likelihood ratio test statistic for the null hypothesis of g = 1 and u = 0, i.e. the premises implied by the REP of infinite horizons and no liquidity constraints. The likelihood ratio test indicates significant deviations from the REP for seven countries. For the remaining four countries, Cyprus, Latvia, the Slovak Republic and Slovenia, the null cannot be rejected at standard levels of significance.12 Given the comparative short sample period we have also estimated pooled regressions – based on different assumptions for the error term – for the countries of our sample. The first regression shows the result for the assumption that the error term is independent of the crosssectional units (countries) and iid normal (that is, the panel is estimated as if it were a crosscountry regression including a constant), the second one shows the results for the assumption of fixed country effects. Both regressions indicate significant deviations from the REP. We also estimated pooled regressions for country groups with similar characteristics (either per-capita GDP or country size). Group 1 comprises the Baltic countries (Estonia, Latvia and Lithuania), group 2 comprises the Czech Republic, Hungary and the Slovak Republic and group 3 Cyprus, Malta and Slovenia.13 The results confirm those obtained for single countries: the REP 7 Data is not (sufficiently) available for disposable income; hence, as suggested by Flavin (1985), we use per-capita income as a proxy. 8 We did not include Bulgaria due to incomplete time series on government debt. 9 Notice that, given the nonstationary nature of the variables included in the analysis, the ARDL(1,2,2) specification admits an error correction representation and a cointegration relationship between Ct, Yt and Gt is assumed to exist. Results concerning the stationarity of the deviations from the long-run equilibrium are available upon request. 10 Results from the unrestricted model implied by Eq. (6) (which does not allow a direct test of the REP) are available upon request. 11 Due to lack of comparability, the estimates of the utility function parameter a are not reported in the table, but are available upon request. 12 We have also estimated the baseline model using quarterly data. Using quarterly data does not change the results substantially. Results are available from the authors upon request. 13 Cyprus, Malta and Slovenia are pooled given their similar per capita GDP and despite the fact that Cyprus and Malta are significantly smaller in country size than Slovenia.
G. Reitschuler / Economic Systems 32 (2008) 119–128
125
Table 1 Parameter estimates for the restricted model
g
d
u
s
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
1.03 (0.72) 0.97*** (0.04) 1.00*** (0.00) 1.00*** (0.00) 1.00*** (0.30) 1.02** (0.54) 1.02*** (0.01) 1.04*** (0.01) 1.01*** (0.00) 1.01*** (0.04) 1.01*** (0.28)
0.96*** (0.00) 2.25 (1.78) 1.72 (1.42) 0.86*** (0.25) 0.99*** (0.28) 0.96*** (0.26) 0.97*** (0.35) 1.08*** (0.04) 1.49*** (0.25) 0.98*** (0.42) 0.98*** (0.05)
0.00 (1.75) 10.34 (1.78) 0.84*** (0.18) 1.28*** (0.18) 0.53 (13.96) 2.24* (1.17) 0.24** (0.12) 1.00*** (0.12) 0.64 (5.31) 1.16 (28.23) 0.08 (3.71)
1.13 0.60 0.96 2.53 0.80 0.49 0.62 0.00 0.05 0.60 1.81
EU-111 EU-112 Group 1 Group 2 Group 3
1.01*** 1.00*** 1.05*** 0.99*** 1.01***
0.95*** (0.05) 1.01*** (0.00) 0.95*** (0.05) 1.04** (0.52) 0.99*** (0.10)
0.91*** (0.01) 1.00*** (0.00) 1.11*** (0.35) 1.77 (2.62) 0.17 (0.16)
(0.00) (0.01) (0.00) (0.00) (0.02)
(4.50) (0.36) (2.15) (2.11) (6.91) (2.64) (0.89) (0.00) (0.74) (21.07) (7.91)
1.11*** (0.14) 1.56*** (0.30) 0.86 (2.35) 0.90 (7.04) 0.69 (0.70)
LRT
R2adj
DW
Obs
0.00 13.84*** 117.95*** 86.83*** 0.00 4.64*** 2.33E+06*** 2.62E+08*** 4.54E+04*** 3.94 0.11
0.94 0.95 0.98 0.98 0.97 0.98 0.96 0.98 0.95 0.98 0.98
2.19 2.15 2.00 2.21 0.88 2.54 1.43 2.52 2.21 2.77 2.07
15 15 14 14 14 14 15 15 15 14 15
1.04E+04*** 9.44E+04*** 17.14*** 0.47 1.65
0.94 0.94 0.96 0.89 0.92
1.33 1.91 1.47 1.73 1.86
161 161 39 39 45
The dependent variable is private consumption in all specifications. Time period: 1990–2006. White heteroskedasticity/serial correlation-corrected standard errors in parenthesis. ***(**)[*] stands for 1% (5%) [10%] significant, LRT is the likelihood-ratio test with the null of Ricardianity. EU-11 is the pooled regression for all countries of the sample with: (1) common intercept, (2) fixed effects; group 1 comprises EST, LVA, LTU, group 2 CZE, HUN, SVK, group 3 CYP, MLT, SVN; DW refers to the Durbin-Watson test statistic; Obs refers to the number of observations.
is rejected for the Baltic States whereas for the other two country groups it cannot be rejected. 3.2. Does debt-neutrality matter? In general, the assumption that consumers internalize the government budget constraint ‘‘imposes formidable requirements on agents’ ability to gather and process information’’ (Lopez et al., 2000). However, the hypothesis that private agents take the government budget constraint into account with regard to their consumption decisions might still be a good approximation of reality. In this respect, the level as well as the increase of government debt may play a decisive role. It is thus probable that individuals develop awareness for the government budget constraint in countries where public debt follows a rapidly rising (or even explosive) path. Then consumers may become aware of future tax implications since a future fiscal adjustment is more likely (Pozzi, 2003, 2006).14 Generally, the evolution of the debt-to-GDP ratio in the NMS shows a somewhat heterogeneous picture. Table 2 gives the development of public debt since 1995 in 3–4-year averages. As can be seen there are mainly three groups of countries: one group (Estonia, Latvia and Slovenia) displays a relatively stable debt evolution, another group (Lithuania and the Slovak Republic) has succeeded in cutting public debt, sometimes even drastically. Yet another group (Cyprus, Malta and Poland) has experienced the opposite, i.e. rapidly rising debt ratios. A second issue to be mentioned is the long-run sustainability of public finances:15 it should be noted that demographic projections are particularly worrisome in the NMS.16 According to the 14
This notion is mentioned, for instance, by Nicoletti (1988, 1992) and Dalamagas (1993). There are different definitions with regard to long-term sustainability of public finances. We use the one of the European Commission (see European Commission, 2006) that defines a sustainable public finance position as to ‘‘whether on the basis of current policies and projected budgetary trends Member States will: (i) meet the government’s intertemporal budget constraint so that the discounted value of future revenues matches the discounted value of future government expenditures and the level of outstanding debt; and, (ii) continue to comply with the budgetary requirements of EMU, and in particular, the Treaty requirement to keep debt levels below the 60 percent of GDP reference value.’’ Regarding the issue of sustainability in transition countries see also Aristovnik and Bercˇicˇ (2007). 16 This is due to the fact that most countries have relatively low fertility rates, a lower life expectancy than EU-15, and a negative migration balance (see European Commission, 2005). 15
G. Reitschuler / Economic Systems 32 (2008) 119–128
126
Table 2 Evolution of Government Debt-to-GDP ratio in the New Member States Country
AVG 1996–1998
AVG 1999–2002
AVG 2003–2006
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
28.1 13.4 6.5 49.4 13.3 51.1 46.1 26.9 17.1 59.1 28.9
61.7 22.3 5.4 55.3 13.1 23.0 58.8 37.7 23.7 47.4 27.6
70.2 30.7 5.0 58.0 13.1 19.6 74.1 43.5 17.2 38.3 29.4
AVG refers to the average value of public debt for the period indicated; data source is the Ameco Database (European Commission).
European Commission (2005), long-run budgetary pressures due to population ageing (as expressed by the average old-age dependency ratio) are projected to more than double in Cyprus, the Czech Republic, Poland and Slovenia and even triple in the Slovak Republic by 2050. These two issues – the development of the debt ratio over the last decade and long-run sustainability of public finances – make the investigation of a model with government debt worthwhile: if, for instance, the evolution of debt becomes unsustainable, then an expansionary policy will have no effect on output (i.e. the fiscal multiplier is zero). On the other hand, countries curbing government debt should be able to reap the laurels of such sustainable fiscal policies sooner or later. The aim of this section is thus to evaluate whether individuals take these developments into account in their consumption decisions. Table 3 shows the following results for the extended model: the likelihood ratio test for the null hypothesis of g = 1 and u = 0 indicates significant deviations from the REP for five countries. For the remaining six countries (Cyprus, the Czech Republic, Malta, Poland, Romania and the Slovak Republic) the null cannot be rejected at standard levels of significance. Again, pooled regressions were estimated due to the fact that the time series for the countries in question which are presented in the last rows of Table 3 are comparatively short. Both regressions for the whole sample indicate significant deviations Table 3 Parameter estimates for the modified restricted model
g
d
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
1.03** (0.49) 1.92 (1.43) 1.00*** (0.00) 0.98*** (0.00) 1.03*** (0.24) 1.04*** (0.05) 1.00*** (0.01) 1.10*** (0.28) 0.99*** (0.01) 1.01** (0.49) 1.01*** (0.00)
EU-111 EU-112 Group 1 Group 2 Group 3
1.07*** 1.06*** 0.99*** 0.99*** 1.00***
(0.02) (0.00) (0.00) (0.00) (0.00)
0.96*** (0.10) 1.75 (1.49) 2.60 (13.63) 0.97*** (0.27) 0.92** (0.42) 0.96 (0.63) 1.13*** (0.10) 1.17** (0.57) 1.52*** (0.58) 0.98*** (0.37) 0.98*** (0.00) 1.08*** (0.03) 1.04*** (0.00) 0.77 (2.49) 0.03 (0.57) 0.88*** (0.04)
u
s
LRT
R2adj
DW
Obs
0.48 (2.62) 0.28 (98.71) 0.78*** (0.42) 0.13 (246.20) 0.99 (2.96) 1.14 (0.78) 0.06 (0.43) 0.08 (6.50) 0.29 (12.27) 0.92 (20.22) 0.62 (1.13)
0.46 (3.73) 2.57 (6.36) 0.79 (2.62) 0.23 (0.77) 7.48 (15.55) 1.06 (1.14) 1.20 (1.21) 4.11 (4.99) 2.96* (1.77) 0.27 (4.49) 1.52 (1.21)
0.18 5.11* 52.39*** 13.39*** 6898.71*** 10.64*** 0.40 0.58 1.06 0.00 6.15**
0.91 0.92 0.98 0.98 0.98 0.98 0.91 0.98 0.96 0.98 0.98
2.20 2.17 1.64 2.27 1.98 2.59 2.13 2.51 2.07 2.57 1.82
11 11 11 11 11 11 11 11 10 11 11
0.70 (0.45) 0.39 (0.25) 0.18 (2.00) 0.20 (0.98) 0.13** (0.06)
2.62*** (0.42) 2.57*** (0.79) 0.78 (22.18) 0.01 (0.56) 1.01 (0.70)
11.73*** 405.26*** 0.04 0.47 45.49***
0.84 0.91 0.97 0.84 0.90
1.41 2.20 1.47 1.25 2.49
118 118 33 33 31
The dependent variable is private consumption in all specifications. Time period: 1995–2006. White heteroskedasticity/serial correlation-corrected standard errors in parenthesis. ***(**)[*] stands for 1% (5%) [10%] significant, LRT is the likelihood-ratio test with the null of Ricardianity. EU-11 is the pooled regression for all countries of the sample with: (1) common intercept, (2) fixed effects; group 1 comprises EST, LVA, LTU, group 2 CZE, HUN, SVK, group 3 CYP, MLT, SVN; DW refers to the Durbin-Watson test statistic; Obs refers to the number of observations.
G. Reitschuler / Economic Systems 32 (2008) 119–128
127
Table 4 Summary of the LR-test results for the model with and without government budget constraint Country
REP model I
REP model II
Debt Evolution
LTS
FS (st)
FS (lt)
Cyprus Czech Republic Estonia Hungary Latvia Lithuania Malta Poland Romania Slovak Republic Slovenia
0.00 13.84*** 117.95*** 86.83*** 0.00 4.64** 2.3E+06*** 2.6E+08*** 4.5E+04*** 3.94 0.11
0.18 5.11* 52.39*** 13.39*** 6898.71*** 10.64*** 0.40 0.58 1.06 0.00 6.15**
(Rapidly) rising Rising Constant Rising Constant (Rapidly) falling (Rapidly) rising (Rapidly) rising Rising-falling (Rapidly) falling Constant
No No (Yes) (Yes) (Yes) (Yes) (Yes) (No) (Yes) No No
n.a. No Yes No Yes Yes n.a. No Yes Yes Yes
n.a. No Yes No Yes Yes n.a. No Yes No No
The numbers refer to the results of the LR-Test, LTS (long-term sustainability) refers to the average dependency ratio as an indicator for fiscal pressures in the future (source: European Commission, 2005); FS(st) [FS(lt)] refers to the evaluation of shortterm [long-term] fiscal sustainability of Aristovnik and Bercˇicˇ (2007); ***(**)[*] stands for 1% (5%) [10%] significant.
from the REP. As in the section before we have estimated pooled regressions for country groups: we cannot reject the REP for the Baltic countries (Estonia, Latvia, Lithuania) and the group with the Czech Republic, Hungary and the Slovak Republic, whereas the REP is rejected for the country group comprising Cyprus, Malta and Slovenia. We now turn to the comparison of the results from the baseline and the extended model (see Table 4). It shows a change in the validity of the REP from Non-Ricardian (in the baseline model) to Ricardian (in the extended model) for the Czech Republic, Malta, Poland, Romania and a change from Ricardian to Non-Ricardian for Latvia and Slovenia. These results are in accordance with the development of debt, long-run budgetary pressures and the evaluation of (short- and long-term) sustainability by Aristovnik and Bercˇicˇ (2007): the Czech Republic, Malta and Poland (which are all Ricardian in the extended model) display rapidly rising debt ratios; Latvia and Slovenia (which are Non-Ricardian in the extended model) show a constant debt ratio. Obviously in the Slovak Republic (which still displays Ricardian behaviour in the extended model) long-run sustainability plays a more prominent role than the rapidly falling debt ratio in the past decade. Interestingly, all five countries that according to the European Commission (2005) will experience substantial age-related expenditures in the future (which would imply rising debt) are Ricardian in the extended model. 4. Conclusions This piece of research presents evidence concerning the validity of the REP for 11 new EU member states. We have investigated two main questions: first, whether the REP has any relevance for these NMS and secondly, if the evolution of debt plays a role with regard to individuals’ consumption decisions. We have investigated these two issues using the model by Leiderman and Razin (1988) and Khalid (1996), extending the basic model by including the government budget constraint. The baseline setting indicates significant deviations from the REP for 7 out of 11 countries (i.e. for all except Cyprus, Latvia, the Slovak Republic and Slovenia). In the extended model we cannot reject the validity of the REP for Cyprus, the Czech Republic, Malta, Poland, Romania and the Slovak Republic. A comparison of the results from the baseline and the extended model shows a change in the validity of the REP (either from Non-Ricardian to Ricardian or vice versa) for the Czech Republic, Latvia, Malta, Poland, Romania and Slovenia. These results are in line with the evolution of government debt and long-run sustainability in those countries: the Czech Republic, Malta and Poland (rapidly rising debt evolution) display a change from Non-Ricardian to Ricardian behaviour and Latvia and Slovenia (having a constant debt evolution) are Non-Ricardian after having been Ricardian in the baseline model. One policy implication for the NMS concerns the situation after EMU entry: one of the main prerequisites for EMU entry is the 3%-deficit criterion which will restrict the room for fiscal policy manoeuvre for some NMS after joining EMU. For other NMS, which already now display a deficit close
128
G. Reitschuler / Economic Systems 32 (2008) 119–128
to balance or even in surplus (such as the Baltic countries), there may be more room for expansive fiscal policy which, however, could be ineffective (with regard to consumption and thus growth) given that the REP holds. Hence, for those countries for which the REP holds, letting their automatic stabilizers play over the cycle and raising growth (potential) via structural reforms may be more appropriate as compared with implementing discretionary fiscal policy measures. Our results suggest that the NMS should closely watch the development of public debt and longrun sustainability of public finances. Clearly, when debt follows a rising or even explosive path the effectivity of fiscal measures will be reduced since individuals will expect future tax implications, thus rendering fiscal policy ineffective. Hence, reducing long-run budgetary pressures may be an instrument to raise the effectiveness of fiscal policy in the respective countries. References Aristovnik, A., 2006. The influence of fiscal policy and private investment on external imbalances in transition economies. Econ. Bus. Rev. 8, 143–166. Aristovnik, A., Bercˇicˇ, B., 2007. Fiscal sustainability in selected transition countries. MPRA Working Paper 122, Munich. Barro, R.J., 1974. Are government bonds net wealth? J. Polit. Econ. 82, 1095–1117. Bailey, M.J., 1971. National Income and the Price Level. McGraw-Hill, New York. Bernheim, D., 1987. Ricardian equivalence: an evaluation of the theory and evidence. In: Fisher, S. (Ed.), NBER Macroeconomics Annual 1987. MIT Press, Cambridge, MA, pp. 263–304. Dalamagas, B., 1993. How efficient is the substitution of debt for taxes in influencing demand? Appl. Econ. 25, 295–303. Darius, R., 2001. Debt neutrality: theory and evidence from developing countries. Appl. Econ. 33, 49–58. European Commission, 2005. Public finances in EMU 2005. European Economy 3, Brussels. European Commission, 2006. Public finances in EMU 2006. European Economy 3, Brussels. Evans, P., 1993. Consumers are not Ricardian: evidence from 19 countries. Econ. Inquiry 31, 534–548. Feldstein, M., 1982. Government deficits and aggregate demand. J. Monet. Econ. 9, 1–20. Fidrmuc, J., 2003. The Feldstein-Horioka puzzle and twin deficits in selected economies. Econ. Plan. 36, 135–152. Flavin, M., 1985. Excess sensitivity of consumption to current income: liquidity constraints or myopia? Can. J. Econ. 18, 117– 136. Gale, W., Orszag, P., 2004. Budget deficits, national saving, and interest rates. Brookings Papers Econ. Activ. 2, 101–210. Giavazzi, F., Pagano, M., 1990. Can severe fiscal contractions be expansionary? Tales of two small European countries. In: Blanchard, O., Fisher, S. (Eds.), NBER Macroeconomics Annual 1990. MIT Press, Cambridge, MA, pp. 75–111. Giavazzi, F., Jappelli, T., Pagano, M., 2000. Searching for non-linear effects of fiscal policy: evidence from industrial and developing countries. Eur. Econ. Rev. 44, 1259–1289. Hall, R.E., 1978. Stochastic implications of the life cycle-permanent income hypothesis: theory and evidence. J. Polit. Econ. 86, 971–987. Haque, N., Montiel, P., 1989. Consumption in developing countries: tests for liquidity constraints and finite horizons. Rev. Econ. Stat. 7, 408–415. Kaadu, H., Uusku¨la, L., 2004. Liquidity constraints and Ricardian equivalence in Estonia. Bank of Estonia Working Paper 2004, 7, Tallinn. Karras, G., 1994. Government spending and private consumption: some international evidence. J. Money Credit Bank. 26, 9–22. Khalid, A.M., 1996. Ricardian equivalence: empirical evidence from developing economies. J. Develop. Econ. 51, 413–432. Leiderman, L., Razin, A., 1988. Testing Ricardian neutrality with an intertemporal stochastic model. J. Money Credit Bank. 20, 1– 21. Lopez, H.J., Schmidt-Hebbel, K., Serve´n, L., 2000. How effective is fiscal policy in raising national saving? Rev. Econ. Stat. 82, 226– 238. Masson, P., Bayoumi, T., Samiei, H., 1995. Saving behaviour in industrial and developing countries. In: IMF (Eds.), Staff studies for the World economic outlook, World economic and financial surveys. IMF, Washington, DC, pp. 1–27. Nicoletti, G., 1988. Private consumption, inflation and the debt neutrality hypothesis: the case of eight OECD countries. OECD Working Paper 50, Paris. Nicoletti, G., 1992. Is tax-discounting stable over time? Oxford Bull. Econ. Stat. 54, 121–144. Pozzi, L., 2003. Tax discounting in a high-debt economy. Oxford Bull. Econ. Stat. 65, 261–282. Pozzi, L., 2006. Ricardian equivalence under imperfect information. J. Public Econ. 90, 2009–2026. Seater, J.J., Mariano, R.S., 1985. New tests of the life cycle and tax discounting hypotheses. J. Monetary Econ. 15, 195–215.