Pierre-Guillaume Méon
email: pgmeon@ulb.ac.be

Keywords: Monetary union, special interests, referendum, voting, abstention Monetary union, special interests, referendum, voting, abstention. JEL classification: D70, F15, F33 Monetary union, referendum, voting, abstention. JEL classification: D70, F15, F33

Voting and turning out for monetary integration: the case of the French referendum on the Maastricht treaty

I. INTRODUCTION

European monetary integration is in many ways an unparalleled event. For the first time in history, a group of democratic countries freely agreed to forego their monetary sovereignty. Even more unprecedented is the fact that some of them chose to ratify the treaty leading to the abandonment of their national currency by referendum. In other words some European countries chose to directly let their people make the decision to replace their national currencies by the euro. Among others, France did so, on 20 September 1992.

The existence of referenda is in general a godsend for the scientific community, because they are not only opportunities for politicians to check the electorate's mindset, but also the closest to a real life-size experiment social scientists can dream of. The results of such polls have therefore attracted a lot of attention from researchers, and economists in particular, 3 That analysis may then cast a new light on three distinct strands of the literature. First, it relates to the general literature on the political economy of monetary integration. Relevant theoretical arguments can be found in [START_REF] Giovannini | Economic and monetary union: What happened? Exploring the political dimension of optimal currency areas[END_REF], [START_REF] Frieden | Invested Interests: the politics of national economic policies in a world of global finance[END_REF][START_REF] Frieden | Invested Interests: the politics of national economic policies in a world of global finance[END_REF], or [START_REF] Eichengreen | The endogeneity of exchange rate regimes[END_REF]. The latter also provided an empirical contribution to that literature by analysing the voting pattern in the 1896 U.S. presidential election on the assumption that its main issue was the participation of the United States in the Gold standard. Second, the present paper is also naturally related to the literature on European integration. It thus extends the existing literature on the support for the process of European integration by focusing on one of its specific aspects. In so doing, it may shed some light on the distributive implications of EMU, thereby possibly providing some advice as to the policy measures likely to ease the process. It also has shorter-term clear political implications after the French and Dutch 2005 rejection of the European Constitutional Treaty, and insofar as more referenda on monetary union are ahead in the UK, Sweden, Denmark, or in new members of the EU. Third, this piece of work is also connected to the research on the theory of voting and abstention. It should in particular help weighing the importance of issue-specific and structural determinants of the turnout rate.

With this end in view, the rest of the paper is organized as follows. The next section describes in which context the referendum was held, and argues that the issue of monetary integration was central to the public debate preceding the poll. Section 3 discusses the theoretical determinants of the approval and abstention rates. Section 4 describes the econometric specification that is employed, as well as the data set on which it is applied.

Section 5 displays and comments the results of our estimations. Section 6 concludes.

II. THE CONTEXT OF THE REFERENDUM

Before turning to the analysis of the determinants of the vote and abstention in the referendum, it is necessary to put it into perspective and to recall the context in which it was held. The importance of the issue of monetary union in the public debate will then appear more clearly.

Some history

The Treaty on European Union, also known as the Maastricht Treaty, is the result of the negotiations that were closed with the forty-sixth European summit, held in Maastricht, Netherlands, on December 9 and 10, 1991. That treaty made provision for the evolution of the then European Economic Community in the political and economic fields. In particular, it 4 paved the way to European monetary union. It was signed on February 7, 1992, but had to be ratified in all member countries by the end of that year to come into full force. In France, François Mitterrand, the French president, opted for a referendum to ratify the treaty, although there was no formal requirement to do so. The poll was planned for September 20, 1992. The official electoral campaign accordingly took place between September 7 and September 19. The campaign's economic context was marked by the EMS crisis that was partly ignited by the uncertainties pertaining to the ERM's future. The last days preceding the vote had proved particularly troubled. Namely, on September 16, the Spanish peseta had been devalued, while the pound Sterling and the Italian lira, that had already left the ERM, had markedly depreciated. In an attempt to defend the French franc, the French central bank had had to raise the daily interest rate to twenty percent, only two days before the poll.

From a political point of view, the Danish referendum had preceded the French one. It had delivered a negative decision, thereby dramatizing the French poll, not to mention that the Danish result was in part responsible in the stress on the EMS. Moreover, the issue of the ratification of the Maastricht Treaty had cut through traditional French political lines.

Namely, only three parties had taken a unanimous stand: on either extremes of the political spectrum, the Communist Party and the National Front firmly opposed ratification, while on the centre former president Valéry Giscard d'Estaing's Union pour la Démocratie Française (UDF), a traditional pro-European party, defended it.

Other parties were divided. Thus, on the right, future president Jacques Chirac's neogaullist Rassemblement Pour la République (RPR) was split between partisans and opponents of a yes-vote. The former were led by future Prime Minister Edouard Balladur, while the latter followed Charles Pasqua and Philippe Séguin. On the left, the Socialist Party, following François Mitterrand, clearly supported the treaty. However, that party's official stance caused some of its members to secede around Jean-Pierre Chevènement, who took advantage of it to become the herald of left-leaning eurosceptics. Finally, the situation was even more complex in the two green parties. One of them, Génération Ecologie, which lost influence soon afterwards, supported the ratification, while the other one took no position at all, as a result of not being able to find an agreement among its leaders. That situation made the results of the referendum particularly unpredictable.

Unsurprisingly, the results proved close, with the treaty being ratified by a hair's breadth.2 More precisely, 51.04% of voters voted "yes" while the opponents of the ratification gathered 48.95% of expressed votes. However, significant heterogeneity was observed among departments. Namely, yes-votes scored highest in Bas-Rhin with 68.59%, while Somme proved to be the most eurosceptic department with only 44.92% of yes-votes. Similarly, abstention rates differed markedly from one department to another. Thus 44.92% of voters did not turnout in Haute Corse whereas abstainers were only 24.23% in Dordogne. This is precisely those heterogeneities that will be exploited in the rest of the paper.

Monetary union as the key issue of the referendum

The central assumption of the present analysis is that voters were influenced, at least partly, by considerations pertaining to monetary union. To be sure, EMU was not the sole provision of the Maastricht treaty, since it also regulated the Community's common foreign policy, and justice. Moreover, non-economic considerations were not absent from voters' preoccupations. Thus, [START_REF] Boissieu | The political economy of French economic policy in the perspective of EMU[END_REF] mention an exit poll according to which the "loss of national sovereignty", "Brussels technocrats", and "German dominance" were the chief motives put forward by voters who claimed they had voted against ratification.

Similarly, partisans of the treaty put forward "peace", the "building of Europe", and the "need to withstand competition with Japan and the United States" as their main motivations.

Nevertheless, it can be convincingly argued that, at least in the French case, the debate over monetary union was at least one of the key issues that shaped the results of the referendum. Thus, [START_REF] Eichenberg | Europeans and the European support for European Community: the dynamics of public support for European integration[END_REF] compare the debate preceding the referendum to a "public seminar" on the relationship between common policies and national economic performance. Moreover, opponents to the treaty concentrated their criticisms on monetary union, presented as an unacceptable loss of sovereignty. This was true of rightoriented leaders, such as Philippe Séguin, as well as of left-leaning leaders, like Jean-Pierre Chevènement.

There is furthermore strong evidence that economic issues related to the process of European integration receive in general considerable attention in the media. Thus, [START_REF] Eichenberg | The political environment and foreign policy opinions: British attitudes toward European integration, 1972-1979[END_REF] observe that the 60% of the news coverage of European integration in Great Britain involved economic issues. While such a statistic is not available for France, one must admit that a glance at the French press reveals a dominance of economic issues in the news related to the EU. More to the point, one must recall that the EMS crisis had started in the summer of 1992 and was going through renewed fury at the time of the campaign for the referendum. That crisis therefore gave monetary issues strong priority in the public debate. Another argument emphasizing the weight of monetary issues in the referendum is the fact that other economic dimensions of the process of European integration were not at stake. Thus, the Maastricht treaty contains no reference to the Common agricultural policy, or to the structural and cohesion funds. The magnitude of intra-European public transfers was therefore to remain unchanged regardless of the result of the referendum. Moreover, the European single market had been completed at the time the treaty was signed, and its existence was not jeopardized. The 1992 referendum can therefore only be compared with great caution to the referenda on EU-membership held in applicant countries and studied by [START_REF] Fidrmuc | Who Favors Enlargement? Determinants of Support for EU Membership in the Candidate Countries' Referenda[END_REF], because their scope was significantly larger. Therefore, beside institutional and symbolic issues, monetary union was by far the most salient issue in the French referendum.

III. FACTORS BOUND TO INFLUENCE THE VOTE AND THE DECISION TO VOTE

The previous section put forward that monetary union played a prominent role in the referendum on the TEU. Understanding the results of that referendum therefore implies determining how that issue affected voters' choice to cast a "yes" or "no" ballot, as well as their decision to participate in the poll. As the analysis is carried out at the level of departments, this ultimately means relating departments' characteristics to the share of 'yes' votes and the participation rate. We must therefore list a set of determinants of voters' attitude towards monetary union that may be measured, or proxied, at the departmental level. In what follows, we focus on the economic implications of monetary union, expressly leaving aside attitudinal factors, such as partisan identifications, political opinions, or general posture towards European integration. We motivate that choice in the first sub-section. The two subsequent sub-sections are devoted to the expected determinants of voting behaviour and abstention, respectively.

A focus on economic characteristics

The first motivation of the choice to refrain from using attitudinal variables is that, as argued above, the role of partisan affiliations was bound to be limited by the fact that most parties were divided on the issue at stake. Second, as Doyle and [START_REF] Fidrmuc | Who Favors Enlargement? Determinants of Support for EU Membership in the Candidate Countries' Referenda[END_REF] argue, the correlation between attitudes and voting behaviour is not very informative, in spite of its usual statistical significance. In the context of our study, observing a correlation between, for F o r P e e r R e v i e w 7 instance, the proportion of respondents of the Eurobarometer survey who declared themselves supporters of European integration in a department and the share of "yes" votes in that department would not reveal much, bar the fact that voters tend to vote according to their opinion. What matters indeed, are the underlying factors that determine voters' opinions.

Furthermore, as [START_REF] Frieden | Invested Interests: the politics of national economic policies in a world of global finance[END_REF] puts it, what we usually refer to as ideology is not necessarily a black box, but something that can be explained by more fundamental factors, where economic considerations play a key part. As an example, he further argues that the same notion of "national pride" is ironically sometimes used in some countries to oppose devaluations, while it is pretext to criticizing the national currency's overvaluation in others.

Third, it can be argued that the causality between attitude and votes is unclear. Thus, Wlezien et al. (1997) observe that causality may well run from vote choice to attitudinal variables. We therefore exclude attitudinal factors from our analysis and focus on economic characteristics in this study. This choice can be further defended on the ground that economic variables are known to impact citizens' opinion or voting behaviour. For instance, [START_REF] Eichenberg | Europeans and the European support for European Community: the dynamics of public support for European integration[END_REF], [START_REF] Gärtner | Who wants euro and why? Economic explanations of public attitudes towards a single European currency[END_REF], or [START_REF] Erlandsson | Determinants of public opinion in Sweden about joining the EMU[END_REF] observe a significant impact of economic evolutions on the support for European integration, as measured by the Eurobarometer survey. Regarding actual referenda, a significant and theoretically sensible influence of economic factors on votes is reported by Weck-Hanneman (1990), [START_REF] Eichengreen | The endogeneity of exchange rate regimes[END_REF]), Vlachos (2004[START_REF] Fidrmuc | Who Favors Enlargement? Determinants of Support for EU Membership in the Candidate Countries' Referenda[END_REF]. Strikingly, whereas Vlachos (2004) and [START_REF] Fidrmuc | Who Favors Enlargement? Determinants of Support for EU Membership in the Candidate Countries' Referenda[END_REF] focus on European membership, only the studies of [START_REF] Eichengreen | The endogeneity of exchange rate regimes[END_REF] and [START_REF] Jonung | The political economy of monetary unification: the Swedish euro referendum of 2003[END_REF] are devoted to the more specific issue of monetary integration as such. However, the former's focus is the late XIX th century United States, while the latter performs no econometric test.

The determinants of the approval rate

As regards the specific factors that may have influenced voters' behaviour in the 1992 referendum, four broad sets of factors are traditionally pointed out in the literature. Those are individual mobility, macroeconomic shocks, openness to European trade, and the sensitivity of voters to inflation. The first two sets of indicators stem directly from [START_REF] Mundell | A theory of optimum currency areas[END_REF] theory of optimum currency areas.

According to that theory, since monetary integration implies foregoing the independence of monetary policy, its cost will accrue chiefly to those areas that are more exposed to macroeconomic shocks. A first set of factors influencing the support for monetary union in a department should therefore be the characteristics that determine its exposure to or departments that may absorb it differently. [START_REF] Mundell | A theory of optimum currency areas[END_REF] underlines the role of the geographic mobility of workers as the main way to absorb those shocks. We should therefore expect more mobile voters to support monetary integration and vote in favour of the Maastricht treaty. At the level of individual voters, mobility takes another dimension, pertaining to the ease with which a worker who is laid off may find another job. For instance, an unskilled worker may find it more difficult to get a new position than a skilled worker.

Overall, this means that the share of yes votes should be greater in departments where voters are more mobile, both geographically and socially.

The third factor that we take into account is openness to European trade. Arguably, monetary union allows a series of efficiency gains due to the reduction of uncertainty and transaction costs in international transactions. The benefits a voter can expect from EMU therefore depend on the volume of his/her transactions with other European countries.

Moreover, Mac [START_REF] Mac Kinnon | Optimum currency areas[END_REF] showed that a greater openness to international trade would lower the cost of giving up the possibility to use the exchange rate to absorb asymmetric shocks by enhancing the flexibility of nominal wages. For those two reasons, we should expect greater openness to European trade to be associated with greater support for the TEU in French department.

The fourth factor should measure the exposure of voters to the cost of inflation. This is due to the specificity of EMU that implies that price stability be the chief objective of the European central bank. Thus Maastricht treaty's famous article 105 stipulates that "the primary objective of the ESCB shall be to maintain price stability". As inflation has many distributive effects, the ESCB's strong official anti-inflationary commitment was bound to raise support from some voters while attracting the opposition of others. More specifically, voters who hold liquid assets should benefit from monetary integration, whereas those whose wealth consists of real assets should not be affected.

The determinants of the abstention rate

Whereas the rational choice model has proved useful when applied to study voters' behaviour in the voting booth, it has run into the "paradox of not voting" when used to understand voters' participation. The rationale of that paradox is that the probability of influencing the result of an election is so close to zero that no voter should find it optimal to vote, however small the cost of doing so may be. This conclusion not only blatantly contradicts empirical observations, but should also prevent any attempt to link participation [START_REF] Dhillon | Economic theories of voter turnout[END_REF] or [START_REF] Feddersen | Rational choice theory and the paradox of not voting[END_REF] have been used by vote theorists to overcome that difficulty.

The first possibility, put forward by [START_REF] Riker | A theory of the calculus of voting[END_REF], is to directly assume that voters get a consumption benefit from voting. This consumption benefit should help compensate at least partly the cost of voting. Therefore, participation should be positive but also depend on the importance of the election in the eyes of voters. Another strand of explanations of turnout behaviour rests on the assumption that some party or faction leaders can spend resources on mobilizing the electorate, as in [START_REF] Shachar | Follow the leader: theory and evidence on political participation[END_REF]. The importance of the vote for those leaders determines the volume of resources spent on that mobilization effort. Finally, group-based ethical models, such as Harsanyi (1977), contend that voters assess the result of a vote not only by the mere result of the vote but also by the number of voters that took part in that vote. One may moreover add that the legitimacy of an election, and most of all of a referendum, is usually measured not only by the margin of the result but also by the rate of participation. A casual glance at the media coverage of the 2004 election of the members of the European Parliament should suffice to convince the sceptical reader of the empirical relevance of this argument.

All those theoretical arguments have the same implication for the issue at hand in this paper. Namely, voters' participation in the referendum should depend on the same factors as those that influence their voting behaviour. Indeed, those theories explain why the cost of voting should usually not deter all voters from turning out. However, the decision to turnout should ultimately depend on each voter's overall assessment of the importance of his/her vote, which in turn depends on the issue at stake. Conversely, each voter's net benefit from voting depends on the cost of voting he/she faces. Therefore, one should complement the set of factors already mentioned by a measure of the cost of voting. Since such a statistic is bound to be very difficult to gauge directly, a solution is to estimate it through each department's average turnout rate in other elections. In the empirical analysis, we therefore add a measure of average abstention in each department to the set of explanatory variables mentioned above.

Another justification of the inclusion of average abstention among explanatory variables stems from the psycho/sociological approaches of the act of voting. Those approaches emphasize the role of structural factors over contextual factors mentioned above in shaping turnout rates. [START_REF] Opp | Why do people vote? The cognitive-illusion proposition and its test[END_REF] for instance argues that the decision to vote is motivated by an overestimation of one's personal influence on the election's outcome, in addition to moral and social incentives. Empirically, those factors may be captured by average abstention. More recent contributions that depart from the assumption of perfect rationality point toward the same direction. For instance, [START_REF] Demichelis | Learning in elections and voter turnout[END_REF] have thus used learning models of behaviour. An interesting feature of their model is that they could generate hysteresis in the turnout rate. Abstention rates should therefore be correlated across elections.

IV. METHODOLOGY AND DATA

Before describing the data on which it was applied, this section presents the general econometric strategy employed in the analysis.

The empirical strategy

The analysis is performed at the level of departments. This allows to use a sufficient number of observations for a meaningful regression analysis, while resulting in a tolerable data constraint. The variables to be explained are the approval rate and the rate of abstention in each department. As those variables are by construction bounded, our regressions are based on a logit model. Accordingly we do not directly use the approval rate nor the rate of abstention as the dependent variable, but transform them in unbounded variables. More precisely, the general specification of our equations, respectively for the share of yes-votes and for the rate of abstention is the following:

logyes i ln (yes votes i / no votes i ) = a 1 + a 2 X i + u i (1a) logabs i ln (abstainers i / registered voters i ) = b 1 + b 2 Y i + v i (1b)
Where X i and Y i are vectors of department i's characteristics. u i and v i are error terms.

As the basic units of the analysis are aggregate observations, we can resort to an OLS method to estimate the logit model. However, the error term is heteroskedastic. To balance that heteroskedasticity phenomenon, [START_REF] Maddala | Limited-dependent and qualitative variables in econometrics[END_REF] proposes to use the minimum chi-square method, where each observation is multiplied by the following weight, in estimation (1a) and (1b) respectively:

wyes i = [ yes votes i × ( 1 yes votes i / total votes i )] ½ ( 2a
)
wabs i = [ abstainers i × ( 1 abstainers i / registered voters i )] ½ (2b) F o r P e e r R e v i e w 11
Moreover, as both equations refer to two successive decisions, namely the decision to vote and the vote itself, observed in the same departments, residuals u i and v i may be correlated. To take advantage of that property, relationships (1a) and (1b) were consequently estimated jointly thanks to the SURE method.

Data

All French mainland departments were used in the computations. Corsican and overseas departments were excluded, because of both their institutional peculiarities and specific data limitations. Overall, this left us with 93 departments for which the data described below could be gathered.

Both left-hand-side variables were computed from the results published in Le Monde, on September 22 nd , 1992. Since that newspaper is an evening paper, the printed results were those announced by the ministry of the Interior, and not simple survey results. Abstention in other elections were obtained from the same source. More precisely, we used the two presidential elections preceding and following the Maastricht referendum, namely the second rounds of the 1988 and 1995 polls. We chose to gauge average abstention using those elections because, in the French system, presidential elections are the only ones that are held at the national level, like referenda. Moreover, the Maastricht referendum occurred approximately halfway between the two presidential elections, which is a desirable feature of those polls. The second set of right-hand-side variables assesses geographic mobility, and was drawn from two sources. The first source is the 1990 census as provided by INSEE. It directly measures the share of the population that has been living in the same department for at least ten years (same department). The second source originates from the French central bank that let us use its local and regional financial statistics dataset. We extracted per capita outstanding mortgages (mortgage) to measure geographical mobility. The interpretation of that variable is that voters with outstanding mortgages have recently acquired their residence, and therefore do not intend to move out in the short run. It should therefore be interpreted as a measure of geographic immobility.

We expect those indicators to be negatively correlated with the approval rate, since more immobile voters should be more exposed to macroeconomic shocks. One may nevertheless contend that those indicators indeed take past mobility into account.

Accordingly, they may as well reflect workers' true mobility as their past fortune of not having had to move because of adverse shocks. If voters form their opinions on the basis of their past experiences, the sign of the relationship between those indicators and the approval rate may then be reversed. The empirical analysis that follows should help establish the relevant interpretation of those variables.

To complement the assessment of the sensitivity of voters to shocks, we supplement geographic mobility by two measures of social mobility. We thus include the rate of longterm unemployment (long-term unemployment), defined as the share of the unemployed who have remained jobless for more than a year, and the share of the population that has no degree (no degree) in the set of explanatory variables. Both variables were drawn from the 1990 census.

Openness to European trade, defined as the ratio of EC imports and exports to total trade (EC-trade), is computed thanks to the database that the French customs put at our disposal. [START_REF] Eichenberg | Europeans and the European support for European Community: the dynamics of public support for European integration[END_REF] used a similar ratio in their cross-country study of the support for European integration. In his study of the Swedish referendum on EU-membership, Vlachos ( 2004) also included such a variable. We expect that ratio to be positively correlated with the approval rate.

To proxy voters' sensitivity to inflation, we add per capita individual liquid savings (liquidities) to our explanatory variables. That statistic comes form the Banque de France's Finally, we complement our set of explanatory variables with two control variables.

The first one is the share of gainful employees in agriculture in each department (agriculture).

The motivation for the inclusion of that variable is that French farmers had regularly voiced their discontent with European integration and the Common agricultural policy. They may therefore have rejected the treaty on the rebound. Moreover, common wisdom about the results of the referendum was that more agricultural departments had tended to vote against ratification of the treaty. However, as [START_REF] Frieden | Invested Interests: the politics of national economic policies in a world of global finance[END_REF] stressed, farmers are also producers of tradable goods, and may therefore benefit from the elimination of intra-EC exchange-rate uncertainty. 3 The second control variable included among the regressors is a dummy variable accounting for the existence of a frontier between the department studied and European country (border). This variable is set equal to one when such a variable exists and otherwise takes the value zero. This variable controls the positive, or negative, effect that the proximity of other Europeans may have on the approval rate in the referendum. One may also view that dummy as complementing the intra-EC trade variable, which, according to the French customs themselves, is not perfectly accurate.

V. RESULTS

This section presents the main findings of our estimations, and provides an assessment of their significance, followed by robustness checks.

Findings

We ran several regressions following the method described above. According to the SURE method, we estimated the relationships by pair. Namely, the relationship for the approval rate and the one for the abstention rate were estimated jointly with the same set of right-hand-side variables. However, a SURE estimation would be equivalent to a simple OLS estimation if the sets of regressors were the same in both relationships. The inclusion of the average abstention rate on the right-hand-side of the abstention relationship in all estimations 3 In line with [START_REF] Frieden | Invested Interests: the politics of national economic policies in a world of global finance[END_REF] argument, stands a survey commissioned by the « Mouvement des citoyens », a private pro-European association [START_REF] Louis | Les Français et la monnaie unique[END_REF]. According to that survey, in 1993, 81% of French farmers declared that they viewed the fluctuations of the French franc vis-à-vis other European currencies as detrimental to their activity. More to the point, 72% of French farmers stated that they were favourable to the adoption of a common currency. (5) and (6), while system (7) limits the set of explanatory variables to those that exhibited a significant coefficient in previous regressions. The figures in parentheses below the estimated coefficients are the t-values. An asterisk means that the variable has a significant influence at the 90% level of confidence (two asterisks: 95%, three asterisks: 99%). R 2 is the coefficient of determination. The figures in parentheses below the estimated coefficients are the t-values. An asterisk means that the variable has a significant influence at the 90% level of confidence (two asterisks: 95%, three asterisks: 99%). R 2 is the coefficient of determination. A first glance at tables (1) to ( 7) suggests a strong relationship between the economic characteristics of departments and the results of the referendum, both in terms of approval and abstention rate. More precisely a sizeable proportion of the explanatory variables used appear significantly in the regressions, and the system of equations' explained variance reaches more than 66%.

That diagnosis applies chiefly to the relationships devoted to the approval rate. Thus, all the core variables, bar the share of intra-EC trade in total trade, enter regression (6) with a significant coefficient. Moreover, those signs remain consistent across specifications, and are not affected by the inclusion of control variables. Furthermore, their significance also remains unaffected by the inclusion of control variables.

A closer look at tables (1) to ( 7) allows to comment the estimated impact of each variable in more detail. Accordingly, the rate of unemployment and the variables measuring social mobility, in other words the share of the population with no degree and the rate of longterm unemployed, exhibit the predicted signs. 4 Namely, the rate of unemployment, the share of the population with no qualification, and the rate of long-term unemployment are all negatively related with the approval rate. Those results are consonant with [START_REF] Jonung | The political economy of monetary unification: the Swedish euro referendum of 2003[END_REF] observation that the probability of voting for EMU in Sweden increased with education and the share of the working age population that were either unemployed, on sick-leave, or on early retirement. The finding that support for monetary integration decreases in departments with a less educated population is also reminiscent of the results of Doyle and [START_REF] Fidrmuc | Who Favors Enlargement? Determinants of Support for EU Membership in the Candidate Countries' Referenda[END_REF] who observe that EU membership is supported by those with favourable and flexible human capital in new EU member countries. 5

Surprisingly, the share of intra-EC trade never appears significantly in the relationship.

Other definitions of openness to European trade, like the per capita European trade ratio, were also used but none performed better in the regressions. Another surprising result is the sign of the coefficient of per capita liquid savings, which is significantly negative. A possible, though admittedly daring, interpretation of this result is that more risk-averse voters may at the same 4 Strikingly, the result on the unemployment rate may seem at odds with the finding of Doyle and Fidrmuc (2004) that higher unemployment was associated with greater approval in referenda on EU accession held in candidate countries. One must recall however that the scope of those referenda was larger than the scope of the Maastricht referendum. In addition, participation in the EU does not imply, at least in the short run, adoption of the euro. 5 Another interpretation can be conceived if one assumes that knowledge about the European Union increases with education. The negative correlation between lack of education and the share of yes votes may then be interpreted as a consequence of [START_REF] Hayo | Knowledge and attitude towards European monetary union[END_REF] finding that support for European integration is correlated with EU knowledge. More interesting are the coefficients of the geographic mobility variables, which are both consistently positive. Those coefficients both suggest that less mobile voters tended to approve the ratification of the Treaty on the European Union. This result points towards the second interpretation of those variables. Namely, instead of being measures of the intrinsic mobility of workers they may be measures of their past exposure to macroeconomic shocks.

Workers who have not had to move out may feel less exposed to shocks. Moreover, workers who feel less likely to have to move out because of an adverse shock will be more likely to buy a house and get a mortgage. Therefore, if voters assess the probability of being hit by an adverse shock on the base of their past exposure, it is reasonable to contend that the approval rate will be higher in departments whose inhabitants seem less mobile at first sight. Our empirical results are consistent with this interpretation.

By the same token, the sign of the coefficient affecting the share of gainful employees in the agricultural sector is not as extraordinary as it may first seem, in spite of popular accounts of the results of the referendum. Indeed, the fact that departments where the share of gainful employees in agriculture is larger tended to vote in favour of the ratification is consistent with the presentation of that sector as a producer of tradable goods. According to [START_REF] Frieden | Invested Interests: the politics of national economic policies in a world of global finance[END_REF] they should therefore support monetary integration, which is consistent with our empirical results.

Finally, the contiguity of the department with another country does not seem to be related to the approval rate. That control variable was therefore ineffective in explaining the approval rate.

As regards the estimation of the relationship of the abstention rate with the characteristics of departments, the results may look more disappointing than for the approval rate. To be precise, fewer explanatory core variables related to monetary integration appear significant. Thus, the share of the population that has been living in the same department for the last ten years, the unemployment rate, and the long-term unemployment rate, which are significant in the approval rate relationships, appear insignificant in the abstention rate relationships. On the other hand contiguity with the border is significantly associated with the rate of abstention, whereas it is insignificant in the other relationship.

However, four variables that appear in the approval rate relationships also appear significantly in the abstention relationships. This is the case of per capita outstanding 18 mortgage and the share of the population with no degree, which both appear significantly negatively correlated with abstention. 6 Finally, both liquid savings and the share of employment in the agricultural sector appear with a positive and significant sign in the relationship.

It must be admitted that finding an economic interpretation for those results may turn out knotty. Ad hoc attitudinal explanations would probably sound more convincing. However, doing so would go beyond the scope of this paper, and we prefer to let the reader find interpretations of his/her own. Nevertheless, the disappointing performance of economic variables in explaining the rate of abstention is indeed not unexpected. The fact that participation is a challenge to the rational theory of voting was emphasized in the previous section. Moreover previous studies, such a Weck- Hanneman (1990), have also shown that the power of economic variables to predict abstention is significantly smaller than their capacity to explain votes themselves. [START_REF] Matsusaka | Voter turnout: how much can we explain?[END_REF] obtained similar results in explaining turnout decision at the individual level using issue-specific variables. Doyle and [START_REF] Fidrmuc | Who Favors Enlargement? Determinants of Support for EU Membership in the Candidate Countries' Referenda[END_REF] also remark that different considerations may drive voting behaviour and participation in referenda on EU-integration.

More to the point, our study provides a positive result, which points out to the secondary role of contextual variables in shaping the turnout rate. Thus, it appears that the average rate of abstention, computed over the two closest national elections surrounding the Maastricht referendum, is overwhelmingly significant. Moreover, removing that variable from the set of explanatory variables results in a spectacular drop in the system of equations' share of explained variance, as the comparison of estimations ( 7) and ( 8) shows. 7 This result contrasts with [START_REF] Matsusaka | Voter turnout: how much can we explain?[END_REF] who observed that past abstention rates, though significantly correlated with present turnout had a very limited explanatory power.8

A possible implication of this finding is that the factors determining abstention are to a great extent structural and must therefore be found outside the scope of the question at stake in the referendum. One may also remark that this finding is consistent with turnout models that are based on learning. The high explanatory power of average abstention rates would thus 19 be the result of some learning process on the part of voters, leading to a strong persistence of abstention.

Robustness checks

A first concern with our results was that they may be driven by the collinearity of some explanatory variables. It can in particular be argued that the unemployment rate, the long-term unemployment rate, and the share of the population that holds no degree should be correlated. To make sure that this is not the case, we ran each estimation anew while dropping in turn the three variables that can be suspected of being collinear, taking estimation (7) as a benchmark. The results of those estimations are displayed in table 5.

The results of those estimations clearly suggest that multicollinearity is not the driving force of our results. Indeed, all coefficients remain significant, the only exception being the share of gainful employees in agriculture, which only fails to be significant in estimation (10) in the equation that explains the approval rate. Moreover, it appears that the magnitude of those coefficients remains stable across specifications. Multicollinearity therefore does not appear to be an issue in our estimations. 9

A second concern was the measurement of geographic mobility. As a key result of our estimations is that past immobility is positively associated with the approval rate, this issue required closer scrutiny. We therefore complemented the set of explanatory variables by an alternative measure of mobility, namely the share of French citizens who had been naturalized, as provided by the 1990 census. This variable was first simply added to the set of explanatory variables. We then dropped insignificant variables. The results of that test are displayed in column (12) of table 5. They show that the share of naturalized Frenchmen was not correlated to the share of yes vote. However, it was significantly correlated with the abstention rate. However, the same estimation reveals that adding the share of French nationals that had been naturalized to the set of explanatory variables has little effect on the coefficients of the variables that already appeared significantly in estimation (7). However, in both equations, the share of the population that has been living in the same department for at least ten years becomes insignificant. On the other hand, the unemployment rate and the longterm unemployment rate now turn out significant in the expression of the abstention rate.

9 It should also be noted that the coefficient of correlation between any two of those variables, although always significant, never exceeds 0.45 (0.27 between unemployment and long-term unemployment, 0.21 between unemployment and the share of population that has no qualification, and 0.44 between long-term unemployment and the share of the population that has no qualification). The figures in parentheses below the estimated coefficients are the t-values. An asterisk means that the variable has a significant influence at the 90% level of confidence (two asterisks: 95%, three asterisks: 99%). R 2 is the coefficient of determination. The figures in parentheses below the estimated coefficients are the t-values. An asterisk means that the variable has a significant influence at the 90% level of confidence (two asterisks: 95%, three asterisks: 99%). R 2 is the coefficient of determination. The finding that when the share of naturalized nationals is introduced in the set of explanatory variables, the share of the population that has been living in the same department for at least ten years becomes insignificant suggests a strong collinearity between them. We subsequently used the former as a substitute for the latter. The results of that estimation are displayed in estimation (13) of table 5. Here again, it appears that the share of naturalized Frenchmen was not significantly associated to the share of yes votes, but that our main results are little affected. The only striking difference with previous estimations is that the rate of long-term unemployment now fails to be significant, but all the other coefficients remain stable.

F

Our final robustness check was to make sure that our results were robust to using subsamples. We therefore ran our estimation on a sample of departments consisting only of those whose number is uneven, which left us with 47 observations. 10 The results of the estimation on that sample is given in estimation ( 14) of table 5. It can be easily verified that the coefficients estimated on that sample are very close to those obtained with all departments.

VI. CONCLUDING REMARKS

The Maastricht treaty was by all accounts a milestone in the history of European integration. Following the Danish "no" to the treaty, the results of the French referendum were of paramount importance. Indeed, a French disapproval might well have brought the process of European integration to a halt. Such a scenario did not materialize but the results were, somewhat unexpectedly, close. This paper's red thread is that the French referendum can be interpreted as a consultation on monetary union, and that our knowledge of the distributive implications of monetary integration can be improved by studying voting behaviour and abstention at the level of departments. We did so by applying a minimum chi-square model to the approval and abstention rates in French mainland departments, thanks to a weighted SURE method. Two chief conclusions emerge from that econometric exercise.

First, it appears that, as expected, the economic characteristics of departments do a good job in predicting voting behaviour. The results are moreover consistent with predictions based on the theory of monetary integration. Thus, it appears that departments where 10 In France, departments are officially referred to either by their name or by their number, which follows the alphabetical order. Number one is for instance Ain and Yonne is 89. The only exceptions are the metropolitan departments whose numbers range from 90 to 95, which were created after the others, and are mainly located in the Paris area. Focusing on uneven-numbered departments is therefore equivalent to a random draw. On the other hand, departments whose residents seem less mobile, and where the share of gainful employees in the agricultural sector was larger, provided more support to the treaty.

Those a priori disconcerting results can be easily explained once one recognizes that past mobility may be the result of a low exposure to macroeconomic adverse shocks, and that farmers are indeed producers of tradable goods, thereby benefiting from a reduction in exchange rate risk.

Second, the results for the participation rate seem more disconnected from the question asked in the referendum. In fact the main variable accounting for differences in participation across departments is the average rate of participation. This finding points toward more fundamental determinants of the participation rate, or toward models based on learning. This is in itself an interesting result. However, this finding also warrants further research, because the significance of average abstention in our regressions can be interpreted as a measure of our ignorance about the deep causes of participation.

  little doubt that French voters were well aware of the monetary implications of the referendum.

  , the same shock may have different consequences if it hits different regions

  . However, several arguments surveyed by

  credit database. It is expected to be positively associated with the approval rate.

  . From system of regressions (1) to (4), the other sets of variables were entered consecutively in the estimations. Control variables are taken into account in systems

  larger share of their savings in liquid assets and fear the unknown implications of monetary union.

  -term unemployment, and the share of the population that does not hold a degree are high tended to oppose ratification.

  . (2004) "Who wants political integration? Evidence from the Swedish EUmembership referendum", Journal of Public Economics, 88 (7-8), 1589-1604.Weck-Hanneman, H. (1990) "Protectionism in direct democracy", Journal ofInstitutional and Theoretical Economics, 146 (3), 389-418. Wlezien, C., M. Franklin, and D. Twiggs (1997) "Economic perceptions and vote choice: disentangling the endogeneity", Political Behavior, 19 (1), 7-17.

Table 1 :

 1 Descriptive statistics the third quarter of 1992 (unemployment), that was kindly put at our disposal by INSEE, the French national institute of statistics.

		Mean	Std. deviation Minimum	Maximum
	Yes	50.178	5.335	41.06	68.59
	Abstention	28.41	2.28	24.23	34.33
	Average abstention	16.56	2.07	10.97	23.19
	Same department	84.62	5.13	70.20	93.40
	Mortgage	19.08	9.03	6.73	75.95
	Unemployment	10.21	2.14	5.80	15.90
	No degree	26.82	3.60	18.07	34.32
	Long-term unemployment	26.68	4.65	16.10	34.60
	EC-trade	0.67	0.11	0.36	0.90
	Liquidities	79.73	31.53	19.63	313.16
	Agriculture	0.081	0.052	0.0002	0.25
	Naturalized Frenchmen	0.029	0.019	0.003	0.079

The first explanatory variable is intended to measure the importance of macroeconomic shocks in each department. That characteristic is assessed by the

Table 2

 2 Determinants of the approval and abstention rates in the referendum

		Weighted SURE estimates
				(1)		(2)		(3)	(4)
			yes	abstention	yes	abstention	yes	abstention	yes	abstention
	Intercept		0.0636	1.63	0.33	1.61	0.31	1.68	0.578	1.55
			(0.21)	(11.51)	(1.13)	(10.7)	(1.04)	(11.09)	(1.79)	(11.32)
				***		***		***	*	***
	Average			0.0325		0.0336		0.035		0.043
	abstention			(8.81)		(8.52)		(8.94)		(12.07)
				***		***		***		***
	Same		0.00464	0.00362	0.02	0.0034	0.0202	0.00305	0.0216	0.00125
	department		(1.32)	(2.66)	(4.63)	(1.73)	(4.62)	(1.58)	(4.96)	(0.72)
				***	***	*	***		***
	Mortgage		0.00929	0.00071	0.00647	0.00084	0.00637	0.00063	0.0137	0.00867
			(4.89)	(0.99)	(3.98)	(1.12)	(3.85)	(0.85)	(3.42)	(5.05)
			***		***		***		***	***
	Unemployment	0.047	0.00145	0.0329	0.00179	0.0339	0.000233	0.0332	0.00267
			(5.38)	(0.44)	(4.29)	(0.48)	(4.16)	(0.06)	(4.15)	(0.80)
			***		***		***		***
	No degree				0.029	0.00201	0.0282	0.00393	0.0253	0.00651
					(5.66)	(0.85)	(5.01)	(1.57)	(4.42)	(2.87)
					***		***		***	***
	Long-term				0.0136	0.00151	0.0137	0.00185	0.012	0.00156
	unemployment			(2.49)	(0.58)	(2.49)	(0.73)	(2.2)	(0.70)
					**		**		**
	EC-trade						0.062	0.152	0.0164	0.0913
							(0.37)	(2.04)	(0.10)	(1.36)
								**
	Liquidities								0.00222	0.00238
									(2.0)	(5.04)
									**	***
	System
	weighted R 2		0.441	0.547	0.559	0.644
	Number	of		93		93		93		93
	observations

Table 3

 3 Determinants of the approval and abstention rates in the referendum

		Weighted SURE estimates
		(5)		(6)	(7)	(8)
		yes	abstention	yes	abstention	yes	abstention	yes	abstention
	Constant	0.621	1.615	0.485	1.472	0.642	1.46	0.568	0.688
		(2.00)	(11.62)	(1.42)	(10.31)	(2.22)	(17.46)	(1.99)	(8.32)
		**	***		***			**	***
	Average		0.0466		0.0474		0.0465
	abstention		(10.91)		(11.49)		(11.73)
			***		** *		***
	Same	0.0227	0.00145	0.0208	0.00077	0.0231		0.0217
	department	(5.40)	(0.86)	(4.48)	(0.43)	(5.74)		(5.48)
		***		***		***		***
	Mortgage	0.0192	0.00724	0.0192	0.00742	0.0191	0.00791	0.0191	0.00526
		(4.44)	(4.11)	(4.41)	(4.39)	(4.54)	(5.14)	(4.55)	(2.16)
		***	***	***	***	***		***	***
	Unemployment	0.027	0.00137	0.0264	0.00091	0.0255		0.0298
		(3.35)	(0.42)	(3.28)	(0.29)	(3.48)		(4.13)
		***		***				***
	No degree	0.0286	0.00773	0.0287	0.00792	0.0286	0.00719	0.0283	0.00332
		(5.08)	(3.40)	(5.10)	(3.64)	(5.37)	(3.66)	(5.32)	(1.07)
		***	***	***	***			***
	Long-term	0.0146	0.00113	0.0133	0.00266	0.0163		0.0132
	unemployment	(2.73)	(0.52)	(2.42)	(1.23)	(3.21)		(2.65)
		***		**				***
	EC-trade	0.0345	0.00768	0.079	0.0261
		(0.21)	(1.18)	(0.47)	(0.40)
	Liquidities	0.00385	0.00194	0.00382	0.00197	0.00382	0.00209	0.00378	0.0016
		(3.17)	(3.95)	(3.15)	(4.20)	(3.21)	(4.81)	(3.19)	(2.31)
		***	***	***	***		***	***	**
	Agriculture	1.244	0.463	1.41	0.671	1.26	0.675	1.169	0.740
		(2.80)	(2.19)	(2.96)	(3.13)	(2.88)	(3.31)	(2.68)	(2.81)
		***	**	***	***	***	***	***	***
	Border			0.0449	0.0513		0.0473		0.0373
				(0.97)	(2.88)		(3.25)		(1.62)
					***		***
	System
	weighted R 2	0.659	0.6748	0.669	0.52
	Number	93			93	93		93
	of observations

Table 5

 5 Determinants of the approval and abstention rates in the referendum

	Weighted SURE estimates
		(12)	(13)	(14)
		yes	abstention	yes	abstention	yes	abstention
	Constant	0.595	1.585	0.776	1.584	1.592	1.529
		(2.05)	(18.75)	(4.92)	(18.75)	(3.80)	(13.46)
		**	***	***	***	***	***
	Average	0.0220	0.0473		0.0470		0.0458
	abstention	(5.45)	(12.71)		(12.61)		(8.44)
		***	***		***		***
	Same					0.0318
	department					(5.82)
						***
	Mortgage	0.0193	0.00755	0.0158	0.00758	0.0250	0.00772
		(4.57)	(5.10)	(3.51)	(5.13)	(3.80)	(3.07)
		***	***	***	***	***	***
	Unemployment	0.0267	0.00685	0.0263	0.00697	0.0259
		(3.48)	(2.23)	(3.21)	(2.28)	(2.44)
		***	**	***	**	**
	No degree	0.0288	0.00950	0.0268	0.00963	0.0104	0.00519
		(5.38)	(4.87)	(4.57)	(4.94)	(1.51)	(1.78)
		***	***	***	***		*
	Long-term	0.014	0.00689		0.00696	0.0268
	unemployment	(2.67)	(3.71)		(3.84)	(3.57)
		***	***		***	***
	Liquidities	0.00385	0.00201	0.00277	0.00202	0.00487	0.00249
		(3.23)	(4.85)	(2.18)	(4.88)	(2.18)	(2.92)
		***	***	**	***	***	***
	Agriculture	1.227	0.815	1.533	0.838	1.212	0.696
		(2.79)	(4.22)	(3.02)	(4.32)	1.63	(2.09)
		***	***	***	***		**
	Border		0.0361	0.126	0.0389		0.0362
			(2.61)	(2.83)	(2.72)		(1.62)
			**	***	***
	Naturalized		1.68		1.819
	Frenchmen		(4.18)		(4.53)
			***		***
	System
	weighted R 2	0.7047	0.6779	0.6957
	Number	93		93		47
	of observations

This was however unexpected at the time, as De[START_REF] Boissieu | The political economy of French economic policy in the perspective of EMU[END_REF] or[START_REF] Franklin | Uncorking the bottle: popular opposition to European unification in the wake of Maastricht[END_REF] recall.

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

[START_REF] Opp | Why do people vote? The cognitive-illusion proposition and its test[END_REF] also observed a positive impact of education on participation.

In addition to being estimated in a SURE model, abstention equations (7) and (8) were estimated on their own. It results that the adjusted R 2 dropped from 66.05% to 11.88% between the two regressions. In other words, dropping average abstention from the set of explanatory variables results in a 54.17 percentage point drop in the explained variance of the abstention rate. Moreover, average abstention explains up to 48% of the variance of abstention observed in the referendum in a simple bivariate regression.

They however contended that using aggregate abstention figures, as we do here, instead of individual observations may raise the share of variance explained by past abstention.

ACKNOWLEDGEMENTS I would like to thank Andrew Austin, Giuseppe Diana, Michel Dietsch, Khalid Sekkat, and an anonymous referee for insightful suggestions, and Bernard Aubry from INSEE for providing unemployment data. Helpful comments by seminar participants at the University Robert Schuman of Strasbourg and at the University Louis Pasteur of Strasbourg are gratefully acknowledged. I am sorry to say that all remaining errors are mine.

D70 -General < D7 -Analysis of Collective Decision-Making < D -Microeconomics, F33 -International Monetary Arrangements and Institutions < F3 -International Finance < F -International Economics, F15


A Confirmatory Analysis of the Unit Root Hypothesis for OECD Consumption-Income Ratios

Keywords: Consumption, Unit Root, Cross-Dependence, Bootstrap Distribution JEL classification: E21, C15, C23 Consumption, Panel Unit Root Testing, Cross-sectional Dependence, Bootstrap Distribution

A Confirmatory Analysis of the Unit Root Hypothesis for OECD Consumption-Income Ratios

Diego Romero-Avila

Introduction

The examination of the presence of stochastic trends in macroeconomic variables has drawn considerable attention among macroeconomists since the seminal work of [START_REF] Nelson | Trends and random walks in macroeconomic time series: some evidence and implications[END_REF]. The distinction between stochastic and deterministic trends in a macroeconomic series is crucial for understanding the nature of shocks.

So if a series contains a unit root, the variable is path-dependent as its current value heavily depends on past levels. In this case, temporary shocks affect the variable permanently as the effect accumulates over time. In contrast, if a series is stationary, temporary shocks can at most have long-lasting but not permanent effects, since the effect vanishes as time elapses.

In this paper, we investigate the existence of a unit root in the consumptionincome ratio -also known as average propensity to consume (APC)-for a sample of 23 OECD countries over the period . The presence of a unit root in APC carries important implications with regard to the empirical validity of some theoretical hypotheses on consumption behaviour and the econometric modelling of consumption functions. As noted by [START_REF] Sarantis | Is the consumption-income ratio stationary? Evidence from panel unit root tests[END_REF], a unit root in APC would be congruent with the Keynesian absolute income hypothesis, [START_REF] Deaton | Involuntary saving through unanticipated inflation[END_REF] involuntary savings theory and the Marxian undercompensating theory. That would support the lack of mean reversion in APC towards a steady state level in response to shocks. In contrast, [START_REF] Duesenberry | Income, Saving and the Theory of Consumer Behavior[END_REF] relative income hypothesis, [START_REF] Friedman | A Theory of the Consumption Function[END_REF] permanent income hypothesis and [START_REF] Ando | The life-cycle hypothesis of saving: Aggregate implications and tests[END_REF] life-cycle hypothesis imply that APC is best described as stationary in the long-run.

As a result, they predict the existence of a long-run equilibrium relationship between consumption and income, thus leading the consumption-income ratio to converge to a steady-state level. In addition, the great-ratios literature predicts the These theoretical implications have not gone untested. Early studies using conventional unit root tests of the augmented Dickey-Fuller (1979, ADF) type provided evidence supporting the presence of a unit root in APC. These include [START_REF] Drobny | An investigation of the long-run properties of aggregate non-durable consumers' expenditure in the United Kingdom[END_REF], [START_REF] Hall | A systems-approach to the relationship between consumption and wealth[END_REF], [START_REF] Molana | The time-series consumption function: error correction, random walk and the steady state[END_REF][START_REF] Horioka | A cointegration analysis of impact of the age structure of the population on the household saving rate in Japan[END_REF], [START_REF] Bjornland | Structural breaks and stochastic trends in macroeconomic variables in Norway[END_REF] and [START_REF] Cook | The nonstationarity of the consumption-income ratio: Evidence from more powerful Dickey-Fuller tests[END_REF]. 2 In contrast, we also find a series of studies providing some evidence in favour of stationarity in APC. These include [START_REF] Campbell | Does Saving anticipate declining Labor Income? An alternative test of the permanent income hypothesis[END_REF], [START_REF] Lettau | Consumption, aggregate wealth, and expected stock returns[END_REF], [START_REF] Slesnick | Are our data relevant to the theory? The case of aggregate consumption[END_REF] and Urgern-Sternberg (1986). More recently, two studies - [START_REF] Sarantis | Is the consumption-income ratio stationary? Evidence from panel unit root tests[END_REF] and [START_REF] Jin | Cointegration of consumption and disposable income: Evidence from twelve OECD countries[END_REF]-have argued that the conflicting evidence obtained in the literature may be the result of the low power of conventional univariate unit root tests when confronted with small data samples and near unit-root processes (see [START_REF] Campbell | Pitfalls and opportunities: What macroeconomists should know about unit roots[END_REF][START_REF] Dejong | The power problems of unit root tests in time series with autoregressive errors[END_REF]. In response, they employ panel unit root tests which exploit the cross-sectional variability of the data, thus rendering important gains in power. [START_REF] Jin | Cointegration of consumption and disposable income: Evidence from twelve OECD countries[END_REF] applies the test of Levin et al. (2002, LLC hereafter) to a sample of twelve OECD countries over the period . He finds evidence of cointegration between nonstationary income and consumption levels, thus implying stationarity in OECD consumption-income ratios. [START_REF] Sarantis | Is the consumption-income ratio stationary? Evidence from panel unit root tests[END_REF] employ the panel unit root tests of Im et al. (2003, IPS hereafter) and Taylor and Sarno (1988). Their evidence from conventional ADF tests and the two panel unit 1 The analysis at the univariate level of the degree of persistence in consumption-income ratios has not been confined to distinguishing between stationarity and a unit root process. One possible alternative comprises the use of fractional integration techniques. Using fractionally based tests, [START_REF] Gil-Alana | Testing of seasonal fractional integration in UK and Japanese consumption and income[END_REF] find that consumption-income ratios for the UK and Japan can be best characterised by seasonal fractional integration with amplitudes varying across frequencies. 2 Within the great-ratios literature, [START_REF] Serletis | International evidence on the long-run implications of the neoclassical growth model[END_REF], [START_REF] Hossain | Long-run implications of neoclassical growth models: Empirical evidence from Australia, New Zealand, South Korea and Taiwan[END_REF], [START_REF] Harvey | How great are the great ratios?[END_REF] provide very little evidence of stationary consumption-income ratios for several industrialised countries. In this paper we re-examine the stochastic properties of OECD consumptionincome ratios. For that purpose, we conduct unit root testing and take two alternative paths to increase statistical power. First, we use the univariate unit root tests of [START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF] which modify conventional unit root tests through a local-to-unity framework in order to render tests with good size and power properties. The use of these tests allows us to be more confident that failures to reject the null of a unit root are not caused by low statistical power, while rejections are not due to size distortions caused by the presence of a large moving average root in the series (see [START_REF] Perron | Useful modifications to some unit root tests with dependent errors and their local asymptotic properties[END_REF]. For confirmatory purposes, we also compute the univariate stationarity test of Kwiatkowski et al. (1992, KPSS hereafter). Secondly, we deploy panel stationarity and unit root tests which increase power by exploiting the cross-sectional variation of the data. In doing so, we take account of a major caveat applying to the panel unit root tests employed by [START_REF] Sarantis | Is the consumption-income ratio stationary? Evidence from panel unit root tests[END_REF] and [START_REF] Jin | Cointegration of consumption and disposable income: Evidence from twelve OECD countries[END_REF], i.e. the failure to control for error cross-sectional dependence which leads the tests to exhibit severe size distortions. Some studies documenting spurious rejections of the null hypothesis from panel tests that assume cross-sectional independence include O'Connell (1998), [START_REF] Maddala | A comparative study of unit root tests with panel data and a new simple test[END_REF], [START_REF] Strauss | Shortfalls of panel unit root testing[END_REF] and [START_REF] Banerjee | Testing for PPP: Should we use panel methods?[END_REF]. Therefore, we employ the stateof-the-art panel unit root tests of [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] and [START_REF] Pesaran | A Simple Panel Unit Root Test in the Presence of Cross-Section Dependence[END_REF] which explicitly allow for error cross-sectional correlation. [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] control for it through modified bootstrap methods while [START_REF] Pesaran | A Simple Panel Unit Root Test in the Presence of Cross-Section Dependence[END_REF] does so by augmenting standard ADF specifications with cross-sectional averages of lagged levels and first-differences of the underlying series. Since most of the existing panel unit root tests -including the ones employed in our analysis-are constructed in a way that rejection of the null hypothesis of joint nonstationarity tells us only that some but not all cross-sectional units are stationary, it is advisable to complement that analysis with panel tests that take joint stationarity as the null hypothesis. Therefore, as noted by [START_REF] Shin | Mean group tests for stationarity in heterogeneous panels[END_REF], by using panel unit root tests jointly with panel stationarity tests, we may be able to draw definitive conclusions about the stochastic properties of the variable under consideration. In fact, rejection in both panel unit root and stationarity tests would indicate the existence of a mixture of stationarity and nonstationarity in the panel, while failure to reject the null in both tests could lead to inconclusive inferences. In addition, there are two cases leading to definitive conclusions: 1) when there is rejection of the null with the panel stationarity test but not with the panel unit root test, which implies that all cross-sectional units contain a unit root, and 2) when there is rejection with the panel unit root test but not with the panel stationarity test, which points to stationarity in all cross-sectional units.

Therefore, to conduct a confirmatory analysis we complement the use of panel unit root tests with the panel stationarity test proposed by [START_REF] Hadri | Testing for stationarity in heterogeneous panel data[END_REF] which is computed as an average of individual KPSS tests. Since the asymptotic distribution of Hadri's test assumes cross-sectional independence, we allow for general forms of cross-sectional dependence by simulating the bootstrap distribution of Hadri's test following [START_REF] Maddala | A comparative study of unit root tests with panel data and a new simple test[END_REF]. Overall, our confirmatory analysis provides clear-cut support for the existence of a unit root in OECD consumption-income ratios over the post-war era.

The remainder of the paper is structured as follows. Section 2 describes the data and the procedures employed in the analysis for the conduct of unit root testing.

Data and Methodology

Data Description

We employ annual data on the consumption-income ratio for 23 OECD countries over the period 1960-2005. The consumption-income ratios are computed as the ratio of total private consumer's expenditure measured at 2000 prices over total GDP at 2000 prices. The ratio is expressed in natural logs. The primary source of the data is the OECD Economic Outlook, No. 78. The list of OECD countries analysed includes: Australia, Austria, Belgium, Canada, Switzerland, Germany, Denmark, Spain, Finland, France, United Kingdom, Greece, Ireland, Iceland, Italy, Japan, Luxembourg, Netherlands, Norway, New Zealand, Portugal, Sweden and United States.

Econometric Methodology

Univariate unit root tests with good size and power

In order to conduct unit root testing at the univariate level, we employ the GLS M class of tests with good size and power proposed by [START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF] For these tests to exhibit good size properties, it is crucial to select the appropriate lag truncation (k) of the ADF specification. For that purpose, [START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF] develop the modified Akaike information criterion (MAIC) which aims at selecting a relatively long lag-length in the presence of a large negative moving average root (thus preventing size distortions) and a short lag-length when that root is not present (thus avoiding unnecessary loss of power). In our application, we take a maximum lag truncation equal to 8. , where i=1,..…,N and t=1,..….,T stand for the number of panel members and time periods, respectively. The standardised statistic is given by: ( )

Panel unit root and stationarity tests with cross-sectional dependence

) ( ) ( i i NT t t Var t E t N = (1)
where

) ( i t E and ) ( i t Var
are the expected value of the mean and variance, respectively. IPS also proposed the Lagrange Multiplier (LM) test statistic, which after normalisation takes the form:

( ) ( ) Pesaran ( 2003) models cross-sectional correlation using a one-factor model given by

) ( ) ( i i NT LM LM Var LM E LM N = (
) ( ) ( i i NT Max Max Var Max E Max N = (3) where = = N i i NT Max N Max 1 1 , the weighted symmetric test ) ( i WS : ( ) ) ( ) ( i i NT WS WS Var WS E WS N = (
it t i it f u + =
, where t f is the unobserved common factor, i is the factor loading coefficient and it is the idiosyncratic error component. Pesaran augments standard ADF specifications with the cross-sectional averages of lagged levels and first-differences of the series in order to eliminate the cross-sectional dependence embodied in t i f . This renders cross-sectionally augmented ADF t-statistics, i.e.

CADF i for each i, which are denoted by ) , ( T N t i

. [START_REF] Pesaran | A Simple Panel Unit Root Test in the Presence of Cross-Section Dependence[END_REF] constructs a 3 See Smith et al. (2004, pp. 165-166) for details on the bootstrap procedure following [START_REF] Maddala | A comparative study of unit root tests with panel data and a new simple test[END_REF] which generates bootstrap innovations through resampling using a block size of 30 and 20,000 replications. The maximum lag order of autocorrelation used to compute the statistics is set at 8. 4 All the five tests take as the null hypothesis the presence of a unit root for all individuals versus the alternative of stationarity for at least one individual unit. for some i. [START_REF] Hadri | Testing for stationarity in heterogeneous panel data[END_REF] computes the panel stationarity test as the average of univariate KPSS tests:

F

= = = N i T t t i i k S T N 1 1 2 , 2 2 1 ˆ , (7)
where

i T t t i i S T = = 1 2 , 2 2
ˆis the univariate KPSS test for individual i, and

= = t j j i t i S 1 , ,
ˆ stands for the partial sum of the estimated OLS residuals from (6).

2 ˆi represents a consistent estimate of the long-run variance of t i, , which allows for serial correlation and heteroskedasticity across the cross-sectional dimension. 5 Equation ( 7) allows for heterogeneity in the estimation of the long-run variances across units, but homogeneity can also be assumed by replacing 2 ˆi in (7) with

= = N i i N 1 2 1 2
ˆ . For the sake of robustness, we compute the test under both assumptions. After standardising the test, we have (

)

) 1 , 0 ( N N LM d k = µ (8)
5 These are obtained non-parametrically using the quadratic spectral kernel with fixed bandwidth. for the specification with trends. 6 The computation of Hadri's statistic requires the individual series to be cross-sectionally independent along with asymptotic normality. Since these assumptions may be overly strong, we will compute the bootstrap distribution of the panel stationarity test following [START_REF] Maddala | A comparative study of unit root tests with panel data and a new simple test[END_REF] to allow for general forms of cross-sectional dependence, thereby correcting for finite-sample bias.

Empirical results

We report in Table 1 the results of univariate unit root tests with good size and power for the specification with a linear trend since we find evidence supporting the statistical significance of the trend coefficient. Our preliminary findings point to the existence of a unit root in OECD consumption-income ratios, as we fail to reject the unit root null even at the 10% significance level for any of the twenty-three countries under analysis. 7 As regards univariate KPSS tests for the specification with a linear trend -see column 8 of Table 1-, we are unable to reject the null of stationarity at the 1% level for any country. Still, we reject the stationarity null at 5% for Austria and Luxembourg, and at 10% also for Australia, Switzerland, Finland, France, Italy, Japan and the Netherlands.

In all, the analysis with univariate unit root and stationarity tests renders mixed evidence regarding the order of integration of OECD consumption-income ratios: while unit root tests point to the presence of a unit root in the series, the KPSS 6 Hadri's statistic must be compared with the upper tail of the standard normal distribution. 7 To conserve space, we do not show the results from the computation of the univariate unit root tests by [START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF] for the specification without a linear trend. Nonetheless, the results remain fairly unchanged since we can only reject the null of a unit root at a marginal 10% level with some tests for Australia, New Zealand and Portugal. These results are available from the author upon request. 11 stationarity test supports mainly the existence of stationarity in the series. This conflicting evidence should not come up as a surprise since it conforms well with that of early studies employing univariate tests. However, the fact that the unit root tests of [START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF] are expected to show good size and power may indicate that failure to reject the unit root null in OECD consumption-income ratios is not caused by low statistical power, but due to the presence of permanent movements in the series.

[Insert Table 1 about here]

All that said, it is widely recognised in the literature that the use of panel unit root and stationarity tests that exploit the cross-sectional variation of the data leads to a much more efficient way to achieve substantial power gains. However, one important caveat is to be considered when conducting panel unit root testing: traditional panel unit root and stationarity tests derived under the assumption of cross-sectional independence are subject to severe size distortions, which leads to spuriously over-reject the null hypothesis. As the international real business cycle literature has demonstrated, there appear to be strong linkages between macroeconomic aggregates -including consumption-among industrialised countries (see [START_REF] Backus | International real business cycles[END_REF][START_REF] Devereux | Realistic cross-country consumption correlations in a two-country, equilibrium, business-cycle model[END_REF]. As a result, we explicitly allow for cross-sectional dependence in the panel stationarity and unit root tests employed in the analysis.

We now proceed to present the results from the powerful unit root tests of [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF], which control for general forms of cross-dependence and finite-sample bias through residual-based bootstrap methods along similar lines to [START_REF] Maddala | A comparative study of unit root tests with panel data and a new simple test[END_REF]. In deriving the empirical distributions of the five statistics tailored to the structure of the cross-sectional correlation of the error and to the sample size of 2, all of the five tests proposed by [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] are unable to reject the unit root null even at the 10% level irrespective of the inclusion of linear trends in the specification. As regards [START_REF] Pesaran | A Simple Panel Unit Root Test in the Presence of Cross-Section Dependence[END_REF] tests, our findings again confirm the above results, as we fail to reject the null of nonstationarity with any of the two tests regardless of the inclusion of linear trends in the specification.

[Insert Table 2 about here]

Table 3 shows the results of the panel stationarity test of [START_REF] Hadri | Testing for stationarity in heterogeneous panel data[END_REF]. Panel A reports the result of the computation of the test under the assumption of crosssectional independence and asymptotic normality, while Panel B reports the bootstrap critical values allowing for general forms of cross-sectional dependence, thereby correcting for finite-sample bias. Under the assumption of cross-sectional independence, Hadri's test strongly rejects the null of stationarity in favour of a unit root irrespective of the assumption of homogeneity or heterogeneity in the computation of the long-run variance. To deal with the issue of cross-dependence, we compute the bootstrap distribution of Hadri's test, which appears to dramatically shift to the right of the upper tail of the standard normal distribution as expected.

But despite this sharp rise in the critical values, we are still able to reject the null at the 2.5% level, thus supporting the existence of a unit root in OECD consumptionincome ratios. This finding is robust to the degree of heterogeneity assumed in the estimation of the long-run variance and to the inclusion of deterministic trends in the specification. All in all, our thorough investigation of the stochastic properties of OECD consumption-income ratios renders consistent evidence supportive of the unit root hypothesis. This occurs in spite of attempting to raise statistical power by employing 1) univariate unit root tests with good size and power and 2) the panel unit root tests of [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] and [START_REF] Pesaran | A Simple Panel Unit Root Test in the Presence of Cross-Section Dependence[END_REF] along with the panel stationarity test of [START_REF] Hadri | Testing for stationarity in heterogeneous panel data[END_REF], which all exploit the cross-sectional variation of the data. The fact that we obtain these results from panel unit root tests which take nonstationarity as the null as well as from the panel KPSS test which takes the null of stationarity, should make us more confident that OECD consumption-income ratios are best described as nonstationary. Our results appear to conform not only with those from early studies employing conventional unit root tests, but also with [START_REF] Sarantis | Is the consumption-income ratio stationary? Evidence from panel unit root tests[END_REF] that employed the panel unit root tests of IPS and [START_REF] Taylor | The behavior of real exchange rates during the post-Bretton Woods period[END_REF].

[Insert

Conclusions

In this paper we have investigated the existence of a unit root in the consumption-income ratio for a sample of 23 OECD countries over the period 1960-2005. For that purpose, we have conducted unit root testing by taking two alternative avenues to raise statistical power. First, we have employed the univariate unit root tests of [START_REF] Ng | Lag length selection and the construction of unit root tests with good size and power[END_REF] which modify conventional unit root tests through GLS-detrending to yield tests with good size and power. For confirmatory purposes, we have also computed univariate KPSS tests. Second, we have deployed the panel unit root tests of [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] and [START_REF] Pesaran | A Simple Panel Unit Root Test in the Presence of Cross-Section Dependence[END_REF] and the panel stationarity test of [START_REF] Hadri | Testing for stationarity in heterogeneous panel data[END_REF], which all increase power by exploiting the crosssectional variation of the data. All of these panel tests take account of the presence of cross-sectional correlation in the error structure of the panel. Overall, our examination of the stochastic properties of OECD consumptionincome ratios has yielded strong evidence for the existence of a unit root in the series. The fact that we obtain these results employing tests that 1) show good power and size properties, 2) allow for cross-sectional dependence, and 3) take nonstationarity and stationarity as null hypotheses, should make us more confident of the existence of stochastic trends in OECD consumption-income ratios. These results appear to be congruent with the Keynesian absolute income hypothesis, Deaton's involuntary savings theory and the Marxian undercompesating theory.

  long-run of stationary investment-income and consumption-income ratios along the balanced growth path. 1

  to the presence of a unit root in APC for 20 OECD countries over the period1955-1994.

  the results of the analysis of the stochastic properties of OECD consumption-income ratios through unit root testing and Section 4 summarises the main findings and then concludes.

  [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] have developed more powerful variants of some commonly used panel unit root tests, thereby allowing for general forms of cross-sectional dependence through bootstrap methods. The first two tests are the standard IPS tests. The t-bar statistic is computed as an average of individual t-statistics from ADF specifications, i.e.

.

  [START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] also develop panel versions of some powerful modifications of univariate

.

  Since all of these tests assume both cross-sectional independence and asymptotic normality,[START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] develop a modified bootstrap procedure to compute p-values of the statistics which are robust to smallsample bias and to cross-sectional dependencies in the data. 3 t , Max and WS reject the null hypothesis for large negative values of the statistic, while LM and Min reject the null for large positive values. 4

  OECD consumption-income ratios, we employ a block size equal to 30 and 20,000 bootstrap replications. The maximum lag-order for individual specifications is set at 8. As reported in Table

Table 3 about here]

	F o r
	P
	e
	e r
	R e v
	i e
	w
	13

Table 2 : Panel unit root test of Smith et al. (2004) and Pesaran (2003). Consumption-income ratio. OECD countries. 1960-2005.

 2 The bootstrap p-values for the five panel unit root tests of[START_REF] Smith | More powerful panel data unit root tests with an application to the mean reversion in real exchange rates[END_REF] are computed employing 20,000 bootstrap replications and defining a block size equal to 30. The maximum lag order is set to 8. The panel unit root tests of[START_REF] Pesaran | A Simple Panel Unit Root Test in the Presence of Cross-Section Dependence[END_REF] are computed with an optimal lag truncation of three. The 1%, 5% and 10% critical values for the CIPS test (CZ test) are -2.36, -2.20 and -2.11 (-2.87, -2.04 and -1.58), for the specification without trends. The 1%, 5% and 10% critical values for the CIPS test (CZ test) are -2.85, -2.71 and -2.63(-2.67, -1.85 and -1.41), for the specification with trends. These critical values are computed for T=50 and N=20. ***, ** and * imply rejection of the null hypothesis at 1%, 5% and 10%, respectively.

		NO TREND	TREND
		Tests	p-value	Tests	p-value
	t	-1.648	0.181	-2.181	0.303
	Max	-0.983	0.385	-1.653	0.525
	LM	4.034	0.140	4.791	0.435
	Min	2.430	0.160	3.558	0.323
	WS	-1.165	0.451	-2.010	0.589
	CIPS	-1.639		-2.412
	CZ	0.601		-0.714
	Notes:

Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK Submitted Manuscript

 [START_REF] Hadri | Testing for stationarity in heterogeneous panel data[END_REF]for the case of homogeneity and heterogeneity in the estimation of the long-run variance, respectively. ***, ** and * imply rejection of the null hypothesis at 1%, 5% and 10%, respectively.