Explaining non-compliance in the Norwegian coastal cod fishery: an application of the multinomial logit

Keywords: Multinomial logit, fishery regulation, non-compliance

Establishing motive is central to the analysis of criminal behaviour. This paper analyses the range of motives for non-compliant behaviour among coastal cod fishermen. A multinomial logit model is employed to analyse the reasons that underlie various motives. The four motives compared in this study are economic, technological and social, and motives caused by bureaucracy and legitimacy problems. The economic motive is found to be most important for coastal fishermen.

However, the cross-sectional data indicate that motives vary with fishing gear, vessel length, and fishermen's age. The results indicate that identifying motives for noncompliant behaviour is not straightforward but is important for reducing the extent of infringements of the regulations. Fishermen who use gill-net, seine and long line, for example, are more often motivated to non-compliance by technical problems and bureaucracy than hand-jig fishers. The study suggests that in order to prevent noncompliance behaviour, the authorities' management policy should be designed to address the problems encountered by different categories of fishers. For example, offences that are committed because of technical problems should be resolved by increasing the involvement of fishermen in the process of drawing up regulations, whereas offences motivated by economic factors should be reduced by increasing inspections.

The study makes two contributions to the literature. First, competing explanations that have their roots in different paradigms are tested for their contribution for explaining non-compliance among fishermen. Secondly, technological problems that create incentives for infringements are explicitly tested for, which to the best of our knowledge is a topic that has not been addressed before. We perform an empirical analysis of the motives for non-compliance, based on fishermen's choices among alternative explanations of why the regulations are violated. A multinomial logit model is used to analyse the responses of the fishers. The objective is to identify ways for the authorities to reduce non-compliance. The following section reviews a number of theories regarding the motivation of non-compliance behaviour. The empirical model used to analyse the motives for non-compliance is outlined in section III. Data and empirical results are presented in sections IV and V. A discussion of the findings and some policy implications for reducing non-compliance are presented in the final section.

II. MOTIVES FOR NON-COMPLIANCE

Several scholars have carried out empirical analyses of non-compliance among fishermen (e.g. Sutinen, Rieser and Gauvin, 1990;[START_REF] Furlong | The deterrent effect of regulatory enforcement in the fishery[END_REF]Sutinen and Kuperan, 1999;[START_REF] Sutinen | Blue water crime: deterrence, legitimacy and compliance in fisheries[END_REF][START_REF] Hatcher | Normative and social influences affecting compliance with fishery regulations[END_REF][START_REF] Hatcher | Further investigations into the factors affecting compliance with U.K. fishing quotas[END_REF][START_REF] Nielsen | Important factors influencing rule compliance in fisheries: lessons from Denmark[END_REF]. A review of the literature shows that several theories are used for explaining non-compliance in fisheries, the theories are founded on different paradigms employed for explaining human behaviour and have origins in research fields of economics [START_REF] Becker | Crime and punishment: an economic approach[END_REF], psychology [START_REF] Kohlberg | Moral stages and moralization, in Moral Development and Behviour[END_REF] , 1990). 2 A downside of superfluous of theories is that it becomes difficult for management authorities to decide which strategy to employ in order to reduce infringements of regulations. Motives for non-compliance among fishermen differs with backgrounds variables, for example, regional differences, differences between fishing gear, vessel size, and attitudes towards regulation, etc. We employ a multinomial logit model to reveal distinctions in motives for non-compliance among fishers. The procedure systemises information on fishermen's explanations for infringement behaviour with backgrounds variables of demography, firm-related factors and attitude factors. Another advantage of the procedure is that explanations for non-compliance founded on theories from fields of economics, sociology and psychology is tested empirically against each other.

Different motives for non-compliance behaviour are outlined in a questionnaire presented for the fishermen, and the individual respondents are asked to choose the motive that is most appropriate for them for explaining non-compliance behaviour (Table 1).

The first motive outlined is based on the premise that economic reasoning is a driving force for non-compliance behaviour. [START_REF] Becker | Crime and punishment: an economic approach[END_REF] suggests that utilitymaximizing individuals might find it optimal to commit a criminal offence when the expected utility from committing the crime exceeds the utility from engaging in legal activity. Economic reasoning is used to explain non-compliance among fishermen [START_REF] Sutinen | The economics of fisheries law enforcement[END_REF][START_REF] Furlong | The deterrent effect of regulatory enforcement in the fishery[END_REF].

The second option build on that technological conditions of different types of fishing gear have implications for fishermen's ability to comply with the regulations, and therefore technological problems as such is explaining the non-compliance behaviour [START_REF] Squires | Public regulation and the structure of production in multiproduct industries: an application to the New England otter trawl industry[END_REF][START_REF] Kirkley | The technology and management of multispecies fisheries[END_REF]. The technology of gill-nets, for example, makes it difficult for the fisherman to comply with by-catch regulations and quota regulations [START_REF] Alam | Sustainable resource use, economic development, and public regulation: the multiproduct gill net fishery of peninsular Malaysia[END_REF]Thunberg, Bresnyan, Adams 1995). This is due to the fact that several species are harvested simultaneously, and the gill-net are an inefficient gear for selective harvesting. Non-compliance with by-catch regulations is also a problem for fishermen using trawl [START_REF] Nielsen | Important factors influencing rule compliance in fisheries: lessons from Denmark[END_REF].

The third is option is that non-compliance by among fellow fishermen is a main motivator for non-compliance of the individual fishermen. The significance of imitating others behaviour is founded on the theory of social behaviour, which emphasises how individual behaviour is influenced by opinions that are formed within a group. In this perspective the opinions of peers has a key influence via the moral context of the actual decisions taken at sea by fishermen (Sutinen andKuperan, 1999, Sutinen andGauvin, 1988).

Finally, the bureaucracy of regulations is a suggested as reason for infringement behaviour. The argument is that fishermen's perceptions of legitimacy and fairness of the regulations have an impact on compliance (Tyler, 1990;[START_REF] Jentoft | Legitimacy and disappointment in fisheries management[END_REF][START_REF] Hønneland | A model of compliance in fisheries: theoretical foundations and practical applications[END_REF][START_REF] Nielsen | An analytical framework for studying: compliance and legitimacy in fisheries management[END_REF]. In this view, fishermen regard regulations as unfair, bureaucratic, and the law and its institutions as inappropriate, and thus as incentives for non-compliance.

The questionnaire form, where the respondents have to choose between alternative explanations for infringement behaviour, has the advantage that it gives the possibility for performing a systematic empirical analysis. The applied outcome choices for non-compliance denote an exhaustive representation of motives for 6 infringement behaviour based on the literature and relevant for the applied case study of the Norwegian coastal cod fishery. 3 A drawback of questionnaire form applied is that it requires a narrowing of definitions of the motives for non-compliance, but this is obtained to the benefit of providing sufficient observations to undertake an empirical analysis.

An important part of the analysis of the motives is to find explanatory factors that discriminate among motivation groups. Additional information on the respondents that can be broken down into demographic factors (age, member of fishery association, position in the firm, etc.), firm-related factors (number of fishing days, number of fishing permits, region, etc.) and attitude factors are collected. Information of value for understanding the underlying mechanisms for non-compliance is obtained by combining knowledge of motives and explanatory factors. These mechanisms are important as a means of suggesting options to reduce noncompliance for different categories of fishermen. Policy instruments that affect economic incentives, for example, might be important as a means of motivating fishermen to comply with economic conditions, whereas conditions that affect legitimacy problems might be the key to reduce non-compliance among fishermen that regard the regulatory climate as too bureaucratic.

III. THE EMPIRICAL MODEL

The respondents' choices of alternative motives are modelled by a multinomial logit model. The respondents chose from J alternative motives (outcomes), which are indexed j=0,…J. The outcomes cannot be ranked in any way, and the multinomial 4 The model for determining the probability of outcome j is:

1) 1 exp( ) Pr ( | ) 1 exp( ) i j ij i J k i k x y j x x = = = +
, where Pr ij (y i = j x i ) denotes the probability that individual i chooses outcome j. x i represents exogenous variables, and j are the parameters to be estimated. The multinomial logit model is based on the assumption of Independence of Irrelevant Alternatives (IIA) meaning that the odds ratio between any two choices is unaffected by any other alternative choice. 5 The IIA assumption is tested by determining whether the odds ratio between each pair of alternatives is impacted, when observations of other alternative choice are eliminated from the estimation [START_REF] Long | Regression Models for Categorical and Limited Dependent Variables[END_REF]. Rejection of the assumption of independence means that biased predictions of probabilities will be obtained by the multinomial logit model.

The impact of individual regressors on the odds ratio is not obtained in (1), but these marginal effects are derived by differentiation:

2) 0 Pr Pr [ Pr ] Pr [ ] J ij ij j ik k ij j k i x = = =

IV. DESCRIPTION OF DATA

A questionnaire is administered to coastal fishermen holding licenses for catching cod, saithe, and haddock in the coastal fishery. Three hundred randomly selected respondents have answered the questionnaire based on telephone interviews. In order to secure a high likehood an honest responses assurance of individual anonymity and The coastal fishermen in Norway are harvesting several species, and their motives for non-compliance may differ, depending on which species is mainly targeted. For example, the probability that a respondent indicates that regulation is bureaucratic might depend on whether the fisherman is referring to the herring fishery or to the cod fishery. This study focuses on the cod fishery, because this fishery is most valuable in terms of income for the coastal fleet. The 245 respondents analysed are those fishermen indicating that cod is the most important species for them, and that regulation of cod has a larger influence on earnings than any other species.

The summary statistics indicate that the majority of fishermen 63.5% (N=154) identify economic motives for cheating, 18% a technical reason, 6.6% the importance of other fishermen's behaviour, and 12.3% that bureaucracy is their reason for non-compliance (Table 2).

The demographic information indicates that the mean age of the respondents is 52 years, and that they are employed as skippers. The firm-related information reveals that most of the vessels involved operate for about 200 days a year, the average vessel length is 12.8 metres, the average crew size is two, and most companies are individually owned.

Four types of fishing gear are used in the coastal cod fishery: gill-net, Danish seine, long line, and hand jig. The summary statistics shows that hand jig fishermen are less at the influence of others as a motive for cheating. Some coastal fishermen also use purse seine gear, but respondents using this gear type target mainly herring, and they are not included.

Four regions of Norway are surveyed: the counties of Finnmark, Troms, Nordland, and the southern region. An interesting observation is that fishermen from Troms and the southern region seem more likely to refer to the bureaucracy motive as an explanation for non-compliance than fishermen from Finnmark and Nordland.

Attitude factors are measured by Likert variables, which reveal the opinions of respondents to regulatory issues. The attitude variables are based on responses on multiple-choice answers of the form: "I agree with the statement", "I partly agree with the statement", "I am neutral to the statement", "I partly disagree with the statement", or "I disagree with the statement", which are scored on a scale of 1 to 5. 6 The attitude variables are designed to into account the respondents' perceptions of different aspects of the regulations and control of the fishery. Among the aspects addressed are attitudes towards regulation of the fishery, personal experience of fishery regulations, perceptions of the fisheries authorities, and attitudes towards cheating in the fishery. The attitudes variables are important for a deeper understanding of the motives for non-compliance.

V. EMPIRICAL RESULTS

A multinomial logit model is employed ( Factors that are not significant at the p=0.10 level are removed. A likelihood ratio (LR) test testing the hypothesis, H 0 : i = 0 for all , is performed. The test statistic is defined as 2(L 1 -L 0 ), where L 0 is the value of the log-likelihood function when the only explanatory variable is the constant term, and L 1 is the log-likelihood value when all explanatory variables are included. The test statistics rejects the H 0hypothesis (H 0 : i = 0) in model, indicating that the variables contribute to explaining the differences in motives for non-compliance. Moreover, the IIA assumption is confirmed, meaning that the odds of any two choices are independent of the other choices.

The model includes three demographic and firm-related factors: fishermen older than 50 years, vessel's length in metres, and fishermen using hand jig fishing gear. The age of the fishermen is not statistically significant, but grouping the data between fisherman younger and older than 50 years of age is found to be significant. With respect to fishing gear, different gear types have been tested, and the data indicate that the hand jig differs from the other gear types. Several other demographic firmrelated factors are tested but none of them contributed to explaining differences in motives for non-compliance. To this end it is interesting to observe that regional distinctions in motives for non-compliance are not statistically demonstrated in the data. Several of the attitude factors are significant in explaining motives for noncompliance, and these factors are included.

The individual effects of the demographic, firm-related, and attitudes variables are obtained by estimating their marginal effects ( A weak indication is found for that fishermen on larger vessels are less likely to identify bureaucracy as their motive for non-compliance (Table 4). The vessel length is, however, statistically insignificant a result that follows due to multicollinearity in the model. Spearman's rank test justifies the significant correlations between vessel length and the following attitude factors: 1) perceived fairness of regulation, 2) perceived probability of a control of papers being conducted, 3) perceived probability that fishermen in general are cheating. Moreover fishermen older than 50 are less likely to identify technical conditions as reason for non-compliance behaviour.

Turning to the attitude factors, the perceived probability that the public managers conduct paper control is perceived being lower among fishermen that are pointing at economic motives for non-compliance compared to fishermen pointing at other motives (Table 4). The result confirms that low risk of being controlled creates an economic motive for non-compliance. It is also interesting that fishermen that point at economic motives for non-compliance have a low trust in that regulation is perceived as fair among fishermen in general. The economic-oriented fishermen also 12 anticipate the lowest rate of infringement among fishermen in general. The fishermen that are motivated by economic factors also have a low trust in that increased involvement of fishermen in the regulatory process will reduce non-compliance. The analysis reveals that fishermen that point at the economic motive for non-compliance are more likely to be influenced by increased control measures than fishermen that point at other motives for non-compliances.

Fishermen that point at technical reasons for non-compliance are most negative towards the discard of useful fish. A relatively positive attitude towards regulation is found among fishermen who emphasise technical motives for non-compliance. They are more likely to believe that the regulations are perceived as fair by fishermen, and that there is a higher probability of their papers being inspected than among fishermen who refer to the economic motive. For the authorities, it is important to realise that fishermen who mention technical grounds for non-compliance have a positive attitude to the involvement of fishermen in the regulation process in order to reduce non-compliance.

The fishermen who identify non-compliance among fellow fishermen as a reason for non-compliance are relatively negative to the idea that the authorities are managing the fisheries properly. They believe that non-compliance among fishermen is more common than is thought by the fishermen who pointing to economic and bureaucratic motives for non-compliance. Finally, about 12% of the fishermen point at the regulations as an explanation for non-compliance. These fishermen mainly use Danish seine, long line and gill net, and have the lowest belief that targeting other species than cod might help to reduce non-compliance. The reason might be that to using these types of fishing gear makes it is difficult to catch other species than cod.

One might expect increasing the engagement of fishermen in the regulatory process 13 to reduce frustration with bureaucracy as a motive for non-compliance, but this does not seem to be case.

VI. DISCUSSION AND POLICY IMPLICATIONS

The paper addresses the influence of various motives for non-compliance among coastal cod fishermen in Norway. The cross-sectional data show that there are various motives for non-compliance among coastal fishermen. The economic motive for non-compliance is important for most fishermen, a finding that is in line with the results of studies performed by [START_REF] Hatcher | Normative and social influences affecting compliance with fishery regulations[END_REF][START_REF] Hatcher | Further investigations into the factors affecting compliance with U.K. fishing quotas[END_REF], [START_REF] Nielsen | An analytical framework for studying: compliance and legitimacy in fisheries management[END_REF][START_REF] Nielsen | Important factors influencing rule compliance in fisheries: lessons from Denmark[END_REF]Sutinen, Rieser andGauvin (1990). Individual economic conditions play an important role in determining the motives that lead to infringement of the regulations. In particular, we find that fishermen who emphasise economic motives believe in a lower likelihood that their papers will be inspected than fishermen who explain non-compliance in terms of non-economic motives. However, increasing fishery control has often been found to be a relatively costly way to achieve compliance [START_REF] Arnason | Costs of fisheries management: the cases of Iceland, Norway and Newfoundland[END_REF]. For this reason, increasing the involvement of fishermen in the regulatory process is often suggested as a means of reducing non-compliance. This study indicates that increased involvement of fishermen is likely to have a low probability of success among economy-oriented fishermen, compared to fishermen who identify technical reasons for noncompliance. The result is interesting because scholars have discussed whether means that affect economic incentives, legitimacy or technical problems should be employed to reduce non-compliance. Our findings suggest that involving fishermen in the regulatory process would reduce non-compliance among some fishermen, whereas economic conditions might be expected to influence the majority of fishermen. The study is based on the premise that fishermen cannot be regarded as a homogeneous group, since harvesting patterns differ among fishermen, depending on regional differences in fishing fields and abundance of fish, differences in harvesting gears, which means that perceptions of regulation and motives for non-compliance also differ among fishers. The study shows that type of fishing gear plays a significant role in explaining differences in motives for non-compliance among coastal fishermen. Empirical studies by [START_REF] Squires | Public regulation and the structure of production in multiproduct industries: an application to the New England otter trawl industry[END_REF], [START_REF] Kirkley | The technology and management of multispecies fisheries[END_REF] show that gear is crucial for determining production conditions and the ability to adjust to regulation. The present study shows that catch technology also plays an important role in explaining differences in motives for infringing the regulations.

More specifically, we find that hand jig fishermen are less likely to suggest that bureaucracy and technical problems motivate non-compliance than fishermen who use Danish seine, gill-net and long line. The use of a particular technology influences fishermen's perception of regulations, in that it appears that hand jig fishermen perceive a higher degree of fairness of regulation than fishermen who use other types of gear. Their somewhat positive attitude to regulation explains the lower percentage of hand jig fishermen who use bureaucracy as their explanation for non-compliance compared with fishermen who use other types of gear. If we wish to understand what causes the differences in motives, it is also important to note that hand jig fishermen have fewer problems in avoiding by-catch than fishermen using Danish seine and long line. 8 Moreover, fishermen who use other gear types than hand jig indicate that periodic (temporal) regulation is the problem, while hand jig fishermen are relatively more concerned with regulation of minimum fish size. 9

It is also interesting to note that our data do not allow us to establish regional differences in motives for non-compliance. We anticipated that the differences in Eighteen percent of the fishermen identify technical problems as the reason for noncompliance. These fishermen mainly use gill net, long line and Danish seine, and regard discard of useful fish as an important regulation issue. Information on how regulative obstacles are perceived by fishermen who use different gears is important if fishermen are to be involved in the regulatory process. The survey indicates that long-line, Danish seine and gill net fishermen suffer from different technical problems; for example, 40% of long-line and 28% of Danish seine fishermen regard bycatch as a major regulation obstacle, whereas mesh size seems to be a problem for gill net fishermen mentioned by 26 % of this group.

Forty percent of the respondents who mention bureaucracy as an explanation for non-compliance stress that periodical quota regulation is an obstacle. These fishermen mainly use Danish seine, long line and gill net. It seems that regulations are linked to a low faith in the possibility that orienting their catch strategy towards other species might reduce non-compliance. This result might follow because these fishermen have limited possibilities of switching their catch strategy. Periodical regulation is regarded as being unnecessarily restrictive. It is also interesting to note that fishermen who regard the regulations as bureaucratic do not believe that increased involvement of fishermen in the regulatory process would alleviate the situation. This attitude might stem from these fishermen's lack of faith in being heard in the management process [START_REF] Jentoft | Legitimacy and disappointment in fisheries management[END_REF]. Finally, it is remarkable that the fishermen who are most sceptical about public control of the fishery are the small minority who identify non-compliance among fellow fishermen as a motive for the non-compliance conducted by them. On second thoughts, however, this result is perhaps not so surprising, in that it suggests that these fishermen would probably recommend greater government control of their fellow fishermen. 2 Sutinen andKuperan (1999), Hønneland (1999) provide an enriched theoretical introduction to the literature applied for explaining non-compliance in fishery.

F

3 The applied outcomes are designed to cover the most likely explanations for non-compliance behaviour among Norwegian costal fishers and for securing accuracy in responses, we have formulated the motives in a wording to be well understood by the fishermen. None of the respondents failed to decide on what motivates noncompliance behaviour.

4 Choices that cannot be ranked in any way are denoted as unordered (see [START_REF] Amemiya | Qualitative response models[END_REF].

5 An example of rejection of the IIA in the literature is obtained in the red bus/blue bus example, where the odds of transportation choice between car and red bus will be affected when transportation with a blue bus is added as an option. The red and blue buses are naturally close substitutes, and therefore the odds between red bus and car will be reduced, when the blue bus is added as an option [START_REF] Mcfadden | Conditional logit analysis of choice behaviour[END_REF]. 6 The Likert variable is used as a device for transforming attitudes with categories like strongly approve, approve, undecided, disapprove, and strongly disapprove into quantitative variable with numeric values between 1 and 5 [START_REF] Likert | A Technique for the Measurement of Attitudes[END_REF]. 7 A referee suggests that a Hausman test might provide a suitable device for testing the exogeneity of the regressors. The procedure builds on the use of exogenous instruments that are correlated with the regressors without being correlated with the regressant. Unfortunately, those instruments are not at hand, which prevents us from performing the suggested test.

8 33% of the fishermen using hand jig mention that avoiding bycatch in the cod fishery is not easy, for seine and long line the percentages are 65% and 53% respectively. 9 The questionnaire also reveals that periodical regulation is regarded as a problem for 29% of fishermen not using hand jig, but is regarded as a problem by only 9 % of the fishermen using hand jig. On the other hand, while 27% of the hand jig fishermen mention that minimum size is an important regulatory problem, this issue is only mentioned by 7 % of other fishers.

  Obeying regulations may thus depend on the technological characteristics of the fishing technology employed.

  a suitable modelling device.

  . The questionnaire is designed in a form and wording that should obtain accurate responses, the accuracy is of particular importance with respect to the choice between different motives for non-compliance. In the study none of the 300 respondents have failed to respond on what motivates noncompliance behaviour.

  technical and bureaucratic explanations for non-compliance than fishermen using other types of gear. The long-line fishermen are less likely to point

  species and differences in fishing fields along the Norwegian coast, a distance of 2000 kilometres, would have produced different motives for non-compliance in different regions. However, the fact that most respondents are operating in the three northernmost counties of Norway (Finnmark, Troms and Nordland) under much the same harvesting conditions might explain the lack of regional differences in the data.10

10

  Eighty-two of the respondents operate in the Counties of Finnmark, Troms and Nordland.

Table 3

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	related factors and attitude factors. The regressors are exogenous with respect to the
	respondent's decision on what motivates non-compliance behaviour. 7
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). Maximum likelihood estimation is employed for estimating the model. The model is based on demographic and firm-

Table 4

 4 The marginal effects indicate that fishermen using hand jig gear are less likely to point to technical and bureaucracy problems as motivations for noncompliance than fishermen using other types of gear. The result is valuable in that it demonstrates that the fishing technology employed influences the motivation for infringement of the regulations. The results indicate that economic incentives should be used to reduce non-compliance among hand jig fishermen. For the fishermen using seine, gill net and long line the authorities should employ means that address economic motivation, but also means that are directed towards motives based on technology and bureaucratic frustration.

	). Several interesting findings

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Table 1 .

 1 Motives for non-compliance with regulations among fishers Question: In your opinion, what is the most important reason for non-compliance with regulations 1. One has to disobey the regulations in order to obtaining a reasonable income from fishing 2. Technical problems make it difficult to comply with the regulations 3. Because other fishermen are cheating 4. Because the regulations are bureaucratic

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Table 2 .

 2 Summary statistics for respondents on the survey for the coastal cod fishery in

	Norway
		Economy	Technical	Others	Bureaucracy	Total
			problem	cheating
		(N=154)	(N=45)	(N=16)	(N=30)	(N=245)
	Demographic
	Average age	52 years	50 years	48 years	53 years	52 years
	Position in firm
	Skipper	98 %	97 %	100 %	100 %	98 %
	Firm-related
	Mean fishing days	200 days	207 days	201 days	196 days	201 days
	Mean vessel length	12.6 m	13.4 m	15.7 m	11.5 m	12.8 m
	Mean crew size	1.9 men	2.3 men	2.6 men	1.8 men	2.0 men
	Corporate form
	Single ownership	77 %	73 %	50 %	83 %	75 %
	Gear			%
	Gill-net	59.1	20.0	8.2	12.7	= 100 %
	Danish seine	55.3	18.4	10.5	15.8	= 100 %
	Long line	60.5	25.6	0.0	14.0	= 100 %
	Hand Jig	77.8	9.3	5.6	7.4	= 100 %
	Region			%
	Finnmark	64.3	19.0	7.1	9.5	= 100 %
	Troms	55.0	20.0	8.3	16.7	= 100 %
	Nordland Southern regions 1)	69.0 58.1	18.0 16.3	5.0 7.0	8.0 18.6	= 100 % = 100 %
	1) Southern regions include the counties: Nord-Trøndelag, Sør-Trøndelag, Møre and
	Romsdal, Sogn og Fjordane, Hordaland, Rogaland, Vest-Agder.

Table 3 .

 3 Estimates made using the multinomial logit model

		Outcome 1	Outcome 2	Outcome 3	Outcome 4
		Economy	Techn. Probl.	Other cheat	Bureaucracy
	Constant		-3.813**	-3.967*	-2.226
			(-2.73)	(-1.90)	(-1.35)
	Age_50		-0.952**	-1.327**	0.402
			(-2.32)	(-1.99)	(0.85)
	Meter		-0.042	0.073	-0.084**
			(-0.95)	(1.21)	(-1.51)
	Hand Jig		-1.342**	-0.323	-1.266**
			(-2.16)	(-0.38)	(-2.09)
	1. Attitude to public		0.136	-0.392	0.134
	control		(0.88)	(-1.63)	(0.80)
	2. Attitude to discard		-0.294**	-0.138	0.299
			(-2.10)	(-0.51)	(1.21)
	3. Fishermen perceive		0.016**	0.032**	0.010
	regulation as fair		(2.44)	(3.07)	(1.43)
	4. Perceived risk of		0.019**	0.018	0.003
	papers being checked		(2.47)	(1.55)	(0.37)
	5. Attitude towards		0.090**	0.099**	0.055
	cheating of fishermen		(3.23)	(3.17)	(1.69)
	6. Attitude towards		0.572**	0.099	0.175
	fishermen's involvement		(2.34)	(0.30)	(0.82)
	7. Harvesting of		0.074	-0.019	-0.471**
	alternative species LR test on all variables 2 (9) 1) = 97.65*	(0.33)	(-0.06)	(-2.47)
	IIA Test	14.41*	9.69*	11.76*	11.79*
	Note. Reference outcome 1 (economy). For outcome 1 the economics are used as a comparison group.
	Note. ** indicates significance at 5% level, and * indicates significance at the 10% level. Figures in
	parenthesis are the t-statistics.
	1) A LR test with H 0 : i = 0.

Table 4 .

 4 Marginal effects Suggestions by an anonymous referee have substantially improved the paper. Communication with Stein ArneRånes is greatly appreciated. The authors wish to acknowledge the financial support of the Research Council of Norway. The views expressed herein are those of the authors and are not to be attributed to the Research Council.

		Outcome 1	Outcome 2	Outcome 3	Outcome 4
		Economy	Technical	Other cheating	Bureaucracy
			problem
	Age_50	0.105	-0.122**	-0.041	0.058
		(1.56)	(-2.29)	(-1.63)	(1.41)
	Meter	0.009	-0.004	0.002	-0.007
		(1.31)	(-0.78)	(1.41)	(-1.50)
	Hand Jig	0.202**	-0.114**	-0.001	-0.086**
		(3.30)	(-2.61)	(-0.07)	(-2.30)
	1. Attitude to public	-0.016	0.016	-0.012*	0.012
	control	(-0.67)	(0.88)	(-1.74)	(0.77)
	2. Attitude to discard	0.008	-0.039**	-0.003	0.034
		(0.32)	(-2.37)	(-0.47)	(1.51)
	3. Fishermen perceive	-0.003**	0.001**	0.0008**	0.0006
	regulation as fair	(-2.67)	(2.16)	(2.43)	(0.93)
	4. Perceived risk of	-0.002**	0.002**	0.0004	-0.00001
	papers being checked	(-3.01)	(2.47)	(1.26)	(-0.02)
	5. Attitude towards	-0.015**	0.009**	0.002**	0.003
	cheating of fishermen	(-3.08)	(2.92)	(1.97)	(1.19)
	6. Attitude towards	-0.074**	0.066**	-0.0001	0.008
	fishermen's involvement	(-2.19)	(2.37)	(0.001)	(0.39)
	7. Harvesting of	0.030	0.016	0.0006	-0.048**
	alternative species	(0.95)	(0.63)	(0.07)	(-2.60)
	Note. ** indicates significance at 5% level, and * indicates significance at the 10% level.
	Figures in parenthesis are t-values.

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

 Sutinen, J.G and J.R Gauvin (1988)


Ana Belén

Maria Dolores Gadea

José María Serrano

Gracia Andía

María Dolores

Gadea Rivas

José María

Serrano Sanz

Marcela Sabaté

Antonio Montañés

José Luis Fernández

< C2 -Econometric Methods: Single Equation Models < C -Mathematical and Quantitative Methods, F15 -Economic Integration < F1 -Trade < F -International Economics, F36 -Financial Aspects of Economic Integration < F3 -International Finance < F -International Economics

Keywords: C22 -Time-Series Models Import function, Export function, Monetary Integration, Sapnish trade JEL Classification: C22, F15, F36 import and export functions, monetary integration, Spanish trade

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Introduction

European monetary integration has obliged the member states to adopt adjustment policies to fulfil the nominal convergence criteria required by Maastricht.

Along with the budgetary constraints on public deficit and debt, there were price requirements, an intense process of inflation control and of foreign exchange stability in the last years. As a consequence of this control over monetary variables, intraeuropean trade flows may have been conditioned by the favourable evolution of prices, one of the two basic determinants of trade. The compulsory macro-economic stability has altered the series of the indexes of real effective exchange rates of the euro countries, especially of those that started out from a more unstable position, and this has possibly caused a change, not only in the evolution of trade, but also in its explanatory bases. In an area that concentrates the majority of its members' trade flows, and in which neither prices nor exchange rates can be used as instruments of economic policy, 21 st century trade may have to be explained in function of new variables.

Both the effect of the real exchange rate on trade and the present trend towards price stability are widely debated topics in recent international economic literature 1 .

With regard to the first, the influence of the real exchange rate on the trade balance, the debate focuses mainly on whether the volatility of the exchange rate significantly affects trade flows 2 . The second, and predominantly within the European framework of the successful economic and monetary integration, the relevant question is whether a greater nominal stability has been achieved by the member states and whether this has encouraged trade 3 . Furthermore, in this context of greater stability, non-price factors are increasingly important in the determination of trade flows 4 . Against this background, perhaps the main contribution of this article is its macroeconomic treatment of the real exchange rate as a determinant of trade in a more stable framework. The case of Spain is special for the analysis of the possible effects of macroeconomic stability on foreign trade because of its significant inflation differential and the intense appreciation of the peseta at the beginning of the nineties. After the European Monetary System (EMS) crisis and the devaluations of 1992 and 1993, Spain reached monetary stability and, in the mid-nineties, began a strict price control that ended with the integration of the peseta into the euro in 1999. The objective of this article is precisely to analyse the possible effects of the nominal convergence achieved in the nineties on Spanish trade. In order to do this, we have estimated the export and import functions paying special attention to the two fundamental variables -income and relative prices-, especially to the latter due to their singular behaviour throughout the period. Our starting hypothesis is that the macroeconomic stability required by the single currency may have altered the explanatory base of the most recent trade flows and exercised relevant effects on present Spanish commerce.

The article is divided into the following four sections. In the first we describe the methodology, sources and data used in the paper. The second is focused on the estimation of the aggregate import and export goods functions. The third deals with illustrating the effects that the crucial years of nominal convergence have had on commerce. Lastly, the fourth presents the principal conclusions of the work.

Methodology, sources and data

The idea of structural change has always been present in Economics, given that the phenomena analysed may be subject to shocks of diverse kinds modifying the theoretical relationships. In consequence, diverse econometric techniques have tried to test the statistical significance of these changes and the stability of the estimated parameters 5 . The methodology used in this paper studies the presence, effects and Carrying out the habitual logarithmic transformation, the lineal functional form allows us to estimate the elasticities. The sample period runs from Spain's joining of the European Economic Community (1986) until the adoption of the euro in 1999. The periodicity of the analysis is quarterly, a temporal perspective which allows us, apart from having a larger number of observations, to approximate the short-run dynamics of the net external demand when faced with changes in its principal determinants. All the variables are used without filtering and in logarithms; their description, source and notation are presented in the statistical Appendix.

The first step was to analyse the integration order of the variables used; secondly, the long and short-run relationships were estimated testing for the existence of cointegration; thirdly, we looked for the existence of instability in the parameters, determining the point where the structural break was produced. And finally, the model was re-estimated incorporating the changes detected, allowing them to affect the parameters of the model or the stochastic characteristics of the series affected.

The analysis of the order of integration of all the series used was carried out using the well-known test of [START_REF] Dickey | Likelihood ratio statistics for autoregressive time series with a unit root[END_REF], DF, and its extension for quarterly data proposed by Hylleberg, Engel, Granger and Yoo (1999), HEGY. According to the results, the null hypothesis of the existence of the unit root cannot be rejected for any of the series used. As for the frequencies different to zero, only the endogenous variable exports of goods seems to present a seasonal component of a stochastic nature. Nevertheless, we have preferred to maintain the gross series, without filtering, because the analysis of the residuals of the cointegration relationship seems to eliminate the problem of seasonality, the series of world income presenting a deterministic seasonal component 6 .

Being integrated variables, the long-run equilibrium relationship is estimated with the methodology proposed by [START_REF] Phillips | Statistical inference in instrumental variables regression with I(1) processes[END_REF] -fully modified estimation (FM)-and with a model of error correction mechanism (ECM) estimated non-lineally (NLLS). The first mitigates the second order biases that are produced in finite samples when estimating cointegration vectors with static regressions, and permits the use of standard asymptotic inference procedures. The second incorporates jointly the short-run dynamics and the process of adjustment to equilibrium, and offers a more efficient estimation of the long-run parameters. The existence of cointegration is tested for by applying the method of [START_REF] Engle | Co-integration and error correction: representation, estimation and testing[END_REF] on the residuals of the regression in the first case, and in the second the test of Banerjee, Dolado y Mestre (1998) on the adjustment parameter.

The possible instability of the parameters is examined using the tests SupF, MeanF y Lc developed by [START_REF] Hansen | Tests for parameter instability in regressions with I(1) processes[END_REF]. The three tests share the same null hypothesis -stability-, but present differences in the alternative. The first, SupF, tests for the existence of a regime shift; the second, MeanF, for the goodness of a model that contains a stable relationship; and lastly, the Lc is appropriate when the probability of parametric variation appears relatively constant throughout the sample, being at the same time a test of cointegration ( : cointegration; : no cointegration).

0

H A H
If these tests suggest non-stability, rejecting the null hypothesis of stability at the habitual levels of confidence, the analysis will be continued using the tests of Gregory and Hansen (1996) that examine the presence of cointegration with structural change. Starting from the initial model: y t =α+β'x t +ε t a fictitious variable is defined

ϕ tτ =0 if t≤[nτ] =1 if t>[nτ]
where τ ∈ (0,1) indicates the position of the unknown point of change.

The model C/S (regime shift) will have the following expression: y t =α 1 +α 2 ϕ tτ +β' 1 x t +β' 2 x t ϕ tτ +ε t so that the break can be produced both in the independent term and in the parameter of the explanatory variables.

From the residuals of the previous regression estimated by ordinary least squares for all the possible break points covering all the sample except the initial and final 15%, the minimum of the following tests of the family of the traditional ADF (augmented Dickey-Fuller) and Z (Phillips, Phillips-Perron) are selected:

ADF * =inf τ∈T ADF(τ) Z α * =inf τ∈T Z α (τ) Z t * =inf τ∈T Z t (τ) F o r P e e r R e v i e

w

Once the break point that minimises the test of the cointegration relationship has been obtained, the results of the estimation of the export and import functions are analysed, checking whether the change affects the whole vector of explanatory variables or only part of them. Finally, we explore the possibility that the series affected by the break may present a change in their integration order, using the tests proposed by Fernández andPeruga (1999, 2005) 8 . The analysis is carried out on the basis of the following regressions:

[ ]

t i t K i i t t t u Y Y D Y + ∆ + - + = ∆ - = - ∑ 1 1 1 1 δ α µ τ t i t K i i t t t u Y Y D Y + ∆ + + = ∆ - = - ∑ 1 1 2 δ α µ τ [ ] t i t K i i t t t t t u Y Y D Y D Y + ∆ + - + + = ∆ - = - - ∑ 1 1 2 1 1 1 δ γ γ µ τ τ D τt = 0 t<[τT] τ є (0,1) T>=[τT]
According to Fernández andPeruga (1999, 2005), it can be considered that the parameter δ is not constant throughout the sample period and, in consequence, the integration order of the stochastic process will change depending on the subsample considered. So for every possible break point the statistics tα 1 , tα 2 , tγ 1 and tγ 2 are calculated. The first is the t statistic associated with the null hypothesis :

α 0 H 1 =0
H 2 =0
in the third regression, or in other words, the existence of unit roots in the first or in the second part of the sample, respectively. Given that a sequential approach is being applied, the two statistics that we take are the supreme and the mean, as suggested by [START_REF] Hansen | Tests for parameter instability in regressions with I(1) processes[END_REF], giving these values the names Sup tα 1, Mean tα 1, Sup tα 2, Mean tα 2, Sup tγ 1, Mean tγ 1, Sup tγ 2 and Mean tγ 2. Following the indications of Zivot and Andrews (1992), the observation in which the supreme is produced is taken as the estimator of the break point. These observations are called TSup tα 1, TSup tα 2, TSup tγ 1 and TSup tγ 2.

Results of the estimation

The export function

The estimated export function is, as in the majority of previous studies, a demand function, and includes, also as in previous studies, the two determinants habitual under a demand approach: income -a variable of scale or level-and prices -a variable of competitiveness 9 . As a variable of scale we have opted finally for the Gross Domestic Product of the OECD countries (gdpoecd), although at first we also specified the function with the imports of the developed block, an area considered relevant in the analysis Spanish exports. As for the competitiveness variable, we introduced the index of real effective exchange rates calculated both with production prices (ree) and with consumer prices, the latter, perhaps, less appropriate to measure the competitive level of a determined nation, and in both cases against the prices of the industrialised countries, accepted as the main competitors of Spanish exports.

The results of the long-run estimation using the Phillips-Hansen method are presented in Table 1. The Gross Domestic Product of the developed countries turns out to be highly significant in the long-run estimation of Spanish exports with a very high elasticity, around 3. Prices, on the other hand, show a lower significance and a much less elastic behaviour (-0,6). A similar result is obtained estimating the long-run relationship with its short-run dynamics using non-lineal techniques -Table 2. In the long run, the coefficient of the variable of scale is also around 3 and the exchange rate reaches an elasticity of (-0,8). In the short run, the variation of the GDP lagged one period is significant -which due to the dynamics of the short-run adjustment takes a negative sign (-1,4), contrary to that presented by the contemporary first difference-and the change in prices now unlagged (-0,5), with the expected sign and an elasticity inferior to that of the long run. At first, then, this would corroborate a more relevant role in export sales for the income variable than for the competitiveness variable. However, the results with respect to the existence of a cointegration relationship are ambiguous, because the ADF-EG test on the residuals of the FM estimation do not permit the rejection of the presence of a unit root, while the estimated parameter of the error correction mechanism is negative and significant, surpassing the critical values tabulated by [START_REF] Banerjee | Error-correction mechanism tests for cointegration in a single-equation framework[END_REF]. The possibility that a structural break is biasing the results of the cointegration tests towards the acceptance of the null hypothesis should be, therefore, taken into account.

Moreover, examining the data for the export of goods between 1986 and 2000 - 3).

The application of the method proposed by Gregory and Hansen (1996) allows us to accept the cointegration hypothesis with a regime shift situated in the first quarter of 1993 (Table 4). The estimation of the coefficients presented in the same table show that the structural break is significant when it affects the independent term and the price elasticity, but not in the parameter that accompanies the income variable.

The results of the estimation of the error correction mechanism model after introducing the structural break in 1993 only in the price variable, given that the parameter β 2 is not significant in the regime shift model, are those expected a priori in the scenario described (Table 5). Around 1993, competitiveness via prices suffered an intense loss of weight becoming an almost irrelevant variable after this date -the real exchange rate parameter passes from being (-1,5) to (-0,1). A peseta tied first to stability because of the nominal convergence criteria, also to a fixed and irrevocable exchange rate with respect to the euro and, finally to its substitution by the single currency, would explain, together with a more and more controlled differential inflation, the scant role of the real exchange rate in the determination of Spanish exports to its European partners in the final years of the century 11 . On the short-run side, the lagged endogenous variable

F o r P e e r R e v i e w
is almost significant with a coefficient around 0,2, the speed of adjustment to the longrun equilibrium decreasing its exclusion from 0,9 to 0,8 but not the explanatory power of the model. The lagged change in income (-1,9), the contemporary variation of prices (-0,5) and the lagged (0,4), as well as the dummy 93(I) introduced after the break point also reach significance.

With respect to the relevance of the other determinant -income-the results do not differ substantially from previous works, being highly significant and giving high coefficients, always superior to one. However, although the values of the coefficientaround 3-are not very different, in none of the previous analyses has such an elastic behaviour of exports with respect to external income been found. These results could be explained by the peculiarities of the period under analysis. Being the years after the integration of Spain in the European Communities and, thus, of modernisation, integration and convergence, a slightly higher value does not seem strange, indicative of specialisation in fields of greater technological content and in products of higher quality 12 .

The import function

In the estimation of imports, again only the demand side is considered. The supply of foreign goods for import is perfectly elastic, and this eliminates the introduction of biases in the values of the parameters in spite of considering only the demand function.

If the country is relatively small so that its demand of imported goods does not affect the determination of the prices on international markets, the supposition of nonsimultaneity between supply and demand is perfectly reasonable. The import of goods can be explained, in the same way as exports, by the two key variables in the determination of foreign trade flows: income and competitiveness, expressing the volume of Spanish imports in function of the real domestic income and the relative prices. To represent Spanish income, we have chosen the GDP, but disaggregated into its principal components -consumption (cons), investment (inv) and exports (export)-, although we tried modelling with the series of final and domestic demands 13 .

Competitiveness, as in sales, was measured against the most developed countries especially in terms of production prices, although we also carried out estimations with consumer prices.

A first approximation to both determinants through the FM estimation, gives the expected role and sign to the scale and price variables as well as passing the cointegration tests (Table 6). On the income side, the maximum elasticity is obtained for consumption (it reaches unity) as opposed to a lower, although highly significant, level for exports (0,7), and for investment (0,1), the least relevant of the three 14 . As for the elasticity of prices, it reaches 0,8, a coefficient that is not maintained in the specification of the ECM where the real exchange rate loses its significance in the long run(Table 7).

The income side experiences minor changes: consumption continues to register the highest elasticity (1,7), now even higher, while the parameters of investment and exports are both very close to 0,4, the former increasing with respect to the FM estimation and the latter decreasing. In the short term, the contemporary variations of the two variables are again significant with similar coefficients once more (0,3 and 0,4, respectively) and practically identical to those of the long term. Nevertheless the lagged investment and exports are not relevant while the variation of consumption is only relevant with a lag. As for prices, here they are significant: the difference in competitiveness presents the correct sign and a value of 0,6.

Looking at the graph of the series of imports and of trade flow evolution, the atypical behaviour is centred in the second half of the period, from 1992-1994 onwards when, in spite of the devaluations of the peseta and its later tendency to stability, the 8). However, the Lc test does not permit us to reject the null hypothesis of stability and cointegration.

For their part, the GH cointegration tests under the structural change also support the same hypothesis. The regimen shift model indicates the point of change in the year 1993, in the second quarter, and again the parameter of the explanatory variable real exchange rate is only altered significantly from that moment on (Table 9) 15 . After the second quarter prices again have little to say compared to income in Spanish trade (the coefficient changes from 1,1 to 0,1). In the NLLS estimation, the change in behaviour of the competitiveness factor is only present in the short-term, long-run prices being non-significant (Table 10) 16 . While the elasticity of the real exchange rate before the break date is 0,8, it falls from that moment on to under 0,2. The behaviour of the other variables considered in the short run is practically unchanged, while in the long run the disaggregated income again is the most important with a coefficient around 1,7 for consumption, 0,3 for investment and 0,4 for exports.

As for the demand determinant, as on the export side, high income elasticities are obtained 17 . Nevertheless, the consumption and the export coefficients (1,6 y 0,4) are inferior to those presented in [START_REF] García | Funciones trimestrales de exportación e importación para la economía española[END_REF] (2,7 and 0,5), the only work in this case with which a comparison can be established, and only higher for investment (0,3 as opposed to 0,2).

14

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

The effects of nominal convergence

The results obtained in the last section, compared to those of previous works, give, as the most important novelty, the existence of a structural break in the relative prices variable, and its loss of explanatory capacity both in the export and import functions. The extension of the period under analysis until the effective implantation of the euro, taking in the process of nominal convergence has, without doubt, favoured the obtaining of this conclusion.

Nevertheless, and with respect to the export trade flows, it is worth pointing out that, in spite of not having been detected or proved by any previous analyses, because they missed out the last years of the nineties -the crucial years of nominal convergence-, some of them do give interesting results for a first approximation to the question [START_REF] Buisán | El sector exterior en España[END_REF], [START_REF] Mauleón | El saldo comercial en el bienio 1993-94: ¿Asistimos a un cambio estructural en el comportamiento de nuestra balanza comercial?[END_REF] and [START_REF] Montañés | Análisis estructural de las importaciones y exportaciones de bienes de la economía española[END_REF]], or, at least, intuitions about the expected behaviour in the near future [START_REF] García | Funciones trimestrales de exportación e importación para la economía española[END_REF]].

For example, [START_REF] Buisán | El sector exterior en España[END_REF] note the limited capacity of the model until 1992 to predict the trade flows of 1993-1994 and the lower impact in the short run of the variations in competitiveness on the modelling of exports until 1995, something also observed in the sales estimations of [START_REF] Mauleón | El saldo comercial en el bienio 1993-94: ¿Asistimos a un cambio estructural en el comportamiento de nuestra balanza comercial?[END_REF] until 1992 and 1994. [START_REF] Montañés | Análisis estructural de las importaciones y exportaciones de bienes de la economía española[END_REF] in their simulation exercises, anticipate a gradual reduction in the deficit of the sub-balance of manufactured goods, even though the detected structural break after European integration meant an increase in imports and a decrease in exports for given values of the explanatory variables; and they refer to the atypical behaviour of all the simulations between 1992 and 1996, especially of manufactured goods, which narrow, extraordinarily, the gap between imports and exports, the latter even surpassing the former, when this was not really so. Lastly,

F o r P e e r R e v i e w
García and Gordo (1998) point out that in the second half of the nineties the relative weight of the competitiveness variables was much reduced, both in exports and in imports, as a consequence of exchange rate stability and of the strong price controls that were imposed in the country during this period. All these indices point to the strength of Spanish sales against the real exchange rate under the requirement of macroeconomic stability, and with a single currency shared with its principal trade partners -exports to the European Union make up about three quarters of the total in 2000.

The results derived from the analysis of the import flow show similar aspects to that of exports. Spanish foreign purchases lose sensitivity to prices after the devaluations of the early nineties, a change contained in the FM estimation, and reflected only in the short run in the ECM, where long-run prices are not significant.

Furthermore, as with exports, other studies -that do not undertake the analysis of trade in the last years of the decade-found anomalous behaviours in the import functions with respect to the determinant exchange rate; to be expected, nevertheless, when in the second half of the nineties, in spite of the exchange rate showing a more stable situation, although less favourable for foreign purchases with the recuperation in the first half of the competitiveness lost at the end of the eighties, imports continue to grow at a more than noteworthy rhythm. So it does not appear that it is prices that explain the increase.

Signs of this appear in [START_REF] Buisán | El sector exterior en España[END_REF], Montañés andSanso (2000, 2002) and [START_REF] García | Funciones trimestrales de exportación e importación para la economía española[END_REF]. The first authors reveal, in the short term, the decrease -in absolute value-of the parameter of the change in prices on adding the last three years 1992-1995 to the estimation, while in the long run they observe a similar behaviour following the Johansen methodology. Montañés andSanso (2000, 2002) long run the variation of the real exchange rate is small. Finally, [START_REF] García | Funciones trimestrales de exportación e importación para la economía española[END_REF] show, for imports and when they consider the disaggregated final demand, that the equations proposed until 1996 do not pass the tests of stability of the parameters during the period 1993-1996, as well as suspecting some loss of explanatory capacity of the variable real exchange rate in the period of exchange stability and price convergence.

Given these results we have opted to study in greater depth the nature of the changes that affect the real effective exchange rate, analysing not only whether it has modified its behaviour in the trade functions, but also the stochastic properties of the series. The results of applying the test for a change in the integration order of Fernández and Peruga (1999,2005) show that, in effect, the series has changed from presenting a unit root in the first part of the sample to being stationary in the second (Table 11). Of all the series analysed, only the ree surpasses the critical values of the tests at the 1% level, the change in the integration order being between the third quarter of 1992 and the first of 1993. These results illustrate that the devaluations that brought about the adjustment of the nominal exchange rate, together with its later stability and the convergence of prices, have modified the character of the series, explaining its peculiar behaviour in the export and import functions.

In consequence, and although the series is I(1) for the complete sample, it is interesting to ask to what extent it is relevant to estimate cointegration relationships, of long-run equilibrium, when one of the variables included lacks a unit root in the second part of the sample, and so of long-run effect. In fact, the variation in the order of integration detected for the index of real effective exchange rate series, which changes from being I(1) until 1992 to I(0) from then on, would imply the incorporation of this variable in the short run after the break, and at levels and differentiated in the first period. In accordance with the above, we have re-estimated the ECM models for exports and imports with NLLS incorporating the variable exchange rate only in the long run in the first sample period, and in the short run at levels and differences discriminating by periods (Tables 12 and13).

With respect to exports the results are not substantially different to those obtained when considering the series as I(1) during the whole period, but they do offer some interesting nuances. Although both income elasticity and the adjustment of the equilibrium relationship hardly change, the exchange rate is significant and has the expected sign during the first period with an elasticity close to one, being significant in the short run in the second period. So the negative and structural influence that the real appreciation of the peseta had during the years after joining the EEC is maintained.

The import function is more interesting. In this case, and if we consider the estimation with the disaggregated income, the exchange rate, although it presents the expected sign, is not significant in the long run during the first period. Neither is it significant in the short run during the second period though it is in the first. Therefore, imports are much less sensitive to the price variable, reacting only to its variations during the first years after 1986 and before its stabilisation.

To sum up, the results of modelling the export and import functions after incorporating the change in the integration order of the real effective exchange rate, corroborate the loss of explanatory power of the price variable for the export and import flows. Likewise, they maintain the higher sensitivity of exports to the price variable, significant in the long run in the first period and in the short run in the second, compared with imports for which only the price variations in the first period are relevant. With respect to income, the elasticity of the import function is invariably higher than that of the export function. intense rhythm of growth even when the economic situation begins to weaken, sustained by consumption. These changes in the trade flows have broken the reiterated trend of narrowing the trade gap that some works noted.

Conclusions

Looking at both the exports and import functions allows us to draw some To sum up, after 1993 we can already notice the effects of the structural break that the euro is provoking on the Spanish foreign sector of the new century, which, from that moment on, operates within a double exchange scenario: fixed with respect to a new trade area that accounts for two thirds of its trade, and floating against the rest. The nominal convergence prior to monetary integration, has modified the behaviour of the Spanish export and import functions, advancing the new elements that are going to rule trade in the future. It would be, nevertheless, too daring to totally discard competition through prices as a trade determinant. The measurement of the real exchange rate on the basis of variables that do not include other cost factors, and the accelerated process of adjustment experienced in the period of fulfilling the requirements of the single currency, make it necessary to be careful. 1986 1987 1988 1989 1990 1991 1992 1993 1994 1995 1996 1997 1998 1999 2000 [START_REF] Hansen | Tests for parameter instability in regressions with I(1) processes[END_REF]. The fully-modified estimation uses a quadratic kernel following his proposal. We also follow the selection method of bandwidth parameter included in the GAUSS program provided by Hansen. Notes: The critical values are tabulated in Gregory and Hansen (1996). T-ratios in brackets in the estimation of coefficients. In the GH test the error standard is in brackets. *significant at the 5% level; ** significant at the 1% level. [START_REF] Bacchetta | Does exchange rate stability increase trade and welfare?[END_REF], [START_REF] Baum | Nonlinear effects of exchange rate volatility on the volume of bilateral exports[END_REF], [START_REF] Lane | External wealth, the trade balance, and the real exchange rate[END_REF], and [START_REF] Grier | Real exchange rate uncertainty and export performance: evidence from 18 countries[END_REF]. 2 Nachane and Ranade (1998) contrast different hypotheses about the effect of the real exchange rate on trade and find a very significant effect. Also Bahmani-Oskooe (2001) and [START_REF] In | The long-run relationship between the ral exchang rate and terms of trade in OECD countries[END_REF] investigate how the exchange rate affects the trade balance and the terms of trade, respectively. Nevertheless, most of empirical studies analyse the impact of the exchange rate on exports [see, among others, [START_REF] Doyle | Exchange rate volatility and Irish-UK trade[END_REF], [START_REF] Morgenroth | Exchange rates and trade: the case of Irish exports to Britain[END_REF], [START_REF] Aristotelous | The impact of the post-1972 floating exchange-rate regime on US exports[END_REF], [START_REF] Nabli | How does exchange rate policy affect manufactured exports in MENA countries[END_REF], [START_REF] Nowak-Lehmann | Different approaches of modelling reaction lags: how do Chilean manufacturing exports to movements of the real exchange rate?[END_REF], Daraisami (2004) and [START_REF] Hassan | Exchange rate volatility and aggregate export growth in Bangladesh[END_REF]], but there are some analyses for imports [see, for example, [START_REF] Kenen | Measuring and analysing the effects of short-term volatility in real exchange rates[END_REF], [START_REF] Koray | Real exchange rate volatility and U.S. bilateral trade: a VAR approach[END_REF], [START_REF] Arize | Does exchange rate volatility affect import flows in G-7 countires? Evidence from cointegration models[END_REF] and Pattichis et al. (2004)].

% GDP X/GDP M/GDP (X-M)/GDP
3 [START_REF] Hu | The volatility impact of the European monetary system on member and non-member currencies, Aggregate imports and expenditure components in the UK: an empirical analysis[END_REF], [START_REF] Couharde | The equilibrium exchange rates of European currencies and the transition to euro[END_REF], [START_REF] Kempa | Excess volatility of real exchange rates in the EMS: some evidence from structural VARs[END_REF], [START_REF] Heimonen | Stationary of the European real exchange rates -evidence from panel data[END_REF]), Sosvilla-Rivero et al. (1999[START_REF] Gil-Alana | The dynamics of the real exchange rates in Europe: a comparative study across countries using fractional integration[END_REF] analyse the evolution of the real exchange rates and the volatility of the European currencies through the different phases of the EMS. Sosvilla-Rivero and Gil-Pareja ( 2004) study the price convergence of the EU countries. On the positive impact of the coordination of stability policies on Spain see Amuedo-Dorantes and Wheeler (2001). 4 There is also a widespread debate on this third consideration in recent literature. [START_REF] Carter | Changing trade patterns in major OECD countries[END_REF] analyse the change in trade patterns in OECD countries. Gourlay andSeaton (2003, 2004) study, together with the influence of the exchange rate and its variation, the relevance of other non-price factors linked to the characteristics of enterprises. [START_REF] Mahdavi | The response of the US export prices to changes in the dollar´s effective exchange rate: further evidence from industry level data[END_REF] for US, Martínez-Zarzoso and [START_REF] Martínez-Zarzoso | The determinants of trade performance: influence of R&D on export flows[END_REF] for Europe and Gil Pareja (2001) for the European automobile market, also introduce technological progress and productivity as key factors. 5 A complete overview on structural breaks can be consulted in [START_REF] Maddala | Unit Roots, Cointegration and Structural Change[END_REF]. 6 See [START_REF] Hylleberg | Seasonal integration and cointegration[END_REF]. The results of these tests are not presented due to limitations of space. 8 An application of this technique can be found in [START_REF] Serrano | Modelling evolving long-run relationships: the linkages between stock markets in Asia[END_REF]. 9 The authors have carried out a thorough review of the literature on Spanish trade functions after the latest works of [START_REF] Escribano | Predicción y análisis de funciones de exportación e importación en España[END_REF] and Montañés andSanso (2000 y 2002). 10 For a more detailed analysis of the evolution of exchange rate and prices during this period, see [START_REF] Gadea Rivas | La peseta en la cultura de la estabilidad[END_REF].

11 Exports to the EMU represent more than 70% of the total exports and nearly 90% of those destined to the developed block. The exchange rate of the euro with respect to other strong currencies on the marketthe dollar or the yen-it will continue to affect trade.

12 Between 1985 and 2000 Machinery and Equipment experiences a six percent increase (from 10.74% to 16.75%). [START_REF] Martín | La integración de España en la Unión Europea[END_REF] and [START_REF] Myro | España en la Unión Europea: etapas y efectos de la integración[END_REF] also point out an increase in the branches of medium technology and the technology-intensive sectors. Furthermore, there has been a considerable effort to adapt product quality to that of the other Community countries. While in 1985 more than half of the intraindustry exports were of lower quality than the imports, in 1995 the proportion had been cut to a little over a third, the increase being especially strong in products of similar quality. See [START_REF] Gordo | Integración económica, comercio intraindustrial y costes de ajuste[END_REF]. 13 However, the results obtained both for the domestic and the final demands were worse than those for the GDP. Nevertheless, we do observe, along the lines of other works, an inferior income elasticity when we use demand. Furthermore, we opted for the disaggregation of the GDP into its components, because a better fit was achieved and to avoid capturing with the aggregated GDP more than just a strict income elasticity, for example a possible trend increase in the external openness of the Spanish economy during the period. [START_REF] García | Funciones trimestrales de exportación e importación para la economía española[END_REF] components, although instability problems in the 1993-1996 period led them finally to select the total GDP. Abbot and Seddighi (1996) for UK imports and [START_REF] Giovannetti | Aggregate imports and expenditure components in the Italia: an econometric analysis[END_REF] for Italian imports also introduced the components of domestic demand with similar results.

  breaks in trade functions, within the framework of the theory of cointegration. Both the exports and imports functions would have the same expression: c represents the corresponding trade flow -export or import; y the vector of variables of income; and p that of the relative prices, or other factors of competitiveness.

  They are multivariate extensions of the tests of[START_REF] Perron | The great crash, the oil price shock, and the unit root hypothesis[END_REF],[START_REF] Perron | Nonstationarity and level shifts with an application to purchasing power parity[END_REF] andZivot and Andrews (1992) which, under the alternative hypothesis, permit the change of the cointegration relationship at an unknown point of the sample period ( : no cointegration;: cointegration with structural change). Of the different modellings proposed by the authors, we have selected the model C/S, regime shift, considering it to be more adequate for the type of relationship analysed 0 from being able to detect cointegration, offers an estimation of the date (τ) at which the structural change takes place. The expressions are as follows:

Graph 1 -

 1 points to the possibility of a structural break in the early nineties -around 1993-, when export sales seem to accentuate their upward trend. Furthermore, the evolution of the economic situation and the development of one of the two basic explanatory variables, the index of real effective exchange rates, strengthen the hypothesis of the existence of a possible break -see Graph 2. From 1993 onwards, not only does Europe join the expansive cycle initiated previously by the USA, but also the Spanish economy starts out on the path towards macroeconomic stability and Monetary overcome the crisis of the European Monetary System that resulted in three devaluations of the peseta, the series of real exchange rates levels off and maintains a new profile -except for the devaluation of 1995-until the fixing of the conversion rates to the euro in May 1998, backed up both by the nominal exchange rate and the strong price convergence that occurred after 1996 10 . The results of applying the tests of parametric instability with the battery of tests of[START_REF] Hansen | Tests for parameter instability in regressions with I(1) processes[END_REF] confirm the intuition of a structural break.. In effect, both the SupF and the MeanF reject the existence of a stable relationship in the export function as does the Lc with the presence of cointegration (Table

  From the previous results two conclusions can be drawn. In the first place, that the relevance of the real exchange rate could have ended before the new conditions defined by the introduction of the single currency. Without exchange rates, foreign exchange restrictions or monetary control, the real economy becomes more important within the Eurozone. However, to flatly affirm that competitiveness via prices has ceased to be relevant, is perhaps a premature and risky conclusion. Firstly, because the index of the real effective exchange rate -measured in production prices-is only a partial measure of competitiveness. It would be desirable to include other measures such as general price indexes or, especially, labour costs. And secondly, because, as our period of analysis finishes with the introduction of the euro, the first indications of what happened after the single currency have been eliminated. After successfully finishing the nominal convergence in 1998, there began a period of increasingly higher differential inflation and of unit labour costs increased by low productivity, which has possibly broken the trend towards macroeconomic stability. We should ask ourselves, therefore, if the change in the order of integration of the real exchange rate is of a permanent character or if it has been only a passing illusion, the fruit of the accelerated efforts to fulfil the criteria of Maastricht 18 .The recent behaviour of the trade balance -and this would be the second conclusion -already points in this direction -Graph 1. Its evolution shows that, after a period of stability, the trade deficit has grown again, driven by an expanding domestic demand, and by rises in prices and costs. The estimated parameters of the functions are coherent with such an evolution. If exports have suffered, without doubt, from international uncertainty, given their high income elasticity, they might have been further affected -given their greater reaction to the real exchange rate-by a new worsening of the macroeconomic conditions. At the same time imports maintain an

Page

  conclusions, not only to understand the Spanish foreign sector in the final years of the last century better, but also as a lesson for the future, because what happened in those years could be a good indication of what is going to happen in the 21 st century, an indicator of the capacity of the Spanish economy to compete in a new scenario without exchange rates. These conclusions could be extended, if later studies corroborate them, to the other member states of the Economic and Monetary Union (EMU), especially those countries on the periphery of Europe, where the efforts to achieve monetary integration have necessarily been -as in Spain-greater.After the two years of devaluations, years detected as a break point in the estimations of the trade functions, the old role of the real exchange rate, one of the basic macroeconomic determinants of foreign trade flows until then, loses its strength in trade specification. Due to the process of nominal convergence initiated in those years, the series of the index of real effective exchange rates changes from being of first order integration until 1992 to being of zero order integration afterwards, thus indicating the structural break produced. As a result, from that moment on it loses most of its capacity, especially in the long term, to determine trade flows, income taking its place.It is, therefore, difficult to explain the trade of the period as a unity and we end up drawing two stages: the first between 1986 and 1992 and the second from 1993 to 2000. In the first, income and competitiveness continue their importance along the lines of previous periods, the strong sensitivity in the long run of exports to a exchange rate 20 appreciation being highlighted; in the second, the variable of scale takes on the leading role in the relationship and takes away importance from prices, transferring the challenge to be competitive to other fronts. This result is nothing more than a consequence of the place that other non-monetary determinants of competitiveness occupy in the explanation of international trade. They are factors such as economies of scale or product differentiation, linked in the last instance to technological progress and recognised by the neo-classical theoretical paradigm as decisive determinants of the volume and the composition of present trade 19 .

  Figure 1: Evolution of exports, imports and trade balance with respect to the GDP

Figure

  Figure 2: Evolution of determinants of foreign trade

  (The number of observations of the sample is 50, from 1987 (I) to 1999 (II). 1986 is excluded from the analysis to avoid the problems of accounting changes produced in Spanish trade flows as a consequence of European integration. Variables and statistical sources: -Endogenous variables: Exports and imports of goods (Dirección General de Previsión y Coyuntura)-x and m -Exogenous variables: Gross Domestical Product OECD (OECD Main Economic Indicators and Quarterly National Accounts)-gdpoecd Spanish Gross Domestical Product at market prices (Contabilidad Nacional Trimestral)-gdpsp Spanish consumption expenditure (CNT)-cons Spanish gross fixed capital formation in capital goods: machinery and equipment (CNT)-inv Spanish goods exports (CNT)-export IREER against the developed countries' production prices (Dirección General de Previsión y Coyuntura)-ree Page 32 of 39 Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

7

  Nevertheless, we have also tried with model C (change of level), model C/T (change of level with trend and model C/S/T (change of regime and trend). The results present slight variations to those finally presented.

  also achieved a better fit when they disaggregated the GDP into itsPage 33 of 39Editorial Office, Dept of Economics, Warwick University, Coventry CV4 7AL, UK

  , S.,Fernández-Rodríguez, F. and Bajo-Rubio, O. (1999): Exchange rate volatility in the EMS before and after the fall, Applied Economics Letters, 6, 717-722.Vogelsang, T.J. and Perron, P. (1998) Additional test for a unit root allowing for a break in the trend function at an unknown time, International EconomicReview, vol. 39, n. 4, 1073Review, vol. 39, n. 4, -1100. . Zivot, E. and Andrews, D.W.K. (1992) Further evidence on the great crash, the oil-price shock, and the unit-root hypothesis, Journal of Business and Economic Statistics 10, 251-270.

  purchase of goods continues its rising trend at a good rate, driven by the strength of economic activity and a rising domestic demand. Again, and as on the export side, there is evidence of a structural break. Particularly, the SupF and MeanF tests again strongly indicate a structural break, especially the former which suggests the existence of a regime shift (Table

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Table 1 . Estimation of the exports of goods, 1987-1999. FM-PH

 1 We use Parzen's window with a truncation parameter=4. The number of lags in the EG cointegration test is carried out according to the[START_REF] Ng | Unit root tests in ARMA models with data dependent methods for the truncation lag[END_REF] method starting with a maximum lag of 8. t-ratios in brackets; *significant at the 5% level; ** significant al the 1% level.

	Constant	-6.64
		(-6.79)
	gdpoecd	3.05
		(35.81)
	ree	-0.61
		(-3.38)
	ADF-EG	-2.83
	Notes:

Table 2 . Estimation of the exports of goods, 1987-1999. NLLS

 2 The critical values of the cointegration test are tabulated in[START_REF] Banerjee | Error-correction mechanism tests for cointegration in a single-equation framework[END_REF]. t-ratios in brackets; *significant at the 5% level; ** significant at the 1% level. Q shows the p-order autocorrelation test of Ljung-Box.

	Short-run dx t-1 dgdpoecd t dgdpoecd t-1 dree t dree t-1 Constant Long-run gdpoecd t-1 ree t-1 ECM Q1 Q2 Q3 R 2 (Aj.) R 2 Notes: Page 24 of 39 0.03 (0.22) 0.41 (0.58) -1.43 -1.29 (-1.92) (-1.80) -0.53 -0.50 (-1.91) (-1.86) 0.36 (1.28) -3.19 -2.73 (-2.42) (-2.35) 2.95 2.92 (25.88) (24.49) -0.78 -0.81 (-3.98) (-3.99) -0.62 -0.56 (-4.23)* (-4.36)** -0.13 -0.13 (-2.56) (-4.51) -0.08 -0.08 (-2.94) (-3.18) -0.25 -0.26 (-7.73) (-9.55) 0.94 0.94 0.95 0.95 Submitted Manuscript

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

Table 3 . Stability parameter test for the goods exports function

 3

		Test	p-valor
	SupF	19.27	0.01
	MeanF	6.51	0.04
	Lc	0.83	0.02
	Notes: The critical values are tabulated in

Table 4 . Gregory-Hansen test of the model C/S (regime change) for the goods exports function

 4 Estimation of the coefficients: y t =α 1 +α 2 ϕ tτ +β 1 gdpoecd t +β 2 gdpoecd t ϕ tτ +γ 1 ree+γ 2 reeϕ tτ +ε t

	α 1 = -4.38	-4.33
	(-2.25)	(-2.30)
	α 2 = -8.24	-8.15
	(-1.89)	(-1.91)
	β 1 = 3.44	3.51
	(5.42)	(13.05)
	β 2 = 0.08
	(0.12)
	γ 1 = -1.44	-1.51
	(-1.83)	(-3.09)
	γ 2 = 1.66	1.72
	(1.59)	(1.90)
	Gregory-Hansen test:
	ADF	-4.30	τ=93(I)
		(0.48)
	Z t	-11.29**	τ=93(I)
		(0.30)
	Z α	-69.08**	τ=93(I)
		(0.30)

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Table 5 . Estimation of the exports of goods with a break in the exchange rate, 1987-1999. NLLS

 5

	Short-run
	dx t-1	0.25	0.25
		(1.59)	(1.63)
	dgdpoecd t	0.40
		(0.58)
	dgdpoecd t-1	-1.91	-1.93	-1.50
		(-2.67)	(-2.74)	(-2.18)
	dree t	-0.55	-0.56	-0.51
		(-2.03)	(-2.05)	(-1.84)
	dree t-1	0.41	0.44
		(1.50)	(1.67)
	D93(I)	-6.28	-6.41	-4.83
		(-2.75)	(-2.85)	(-2.35)
	Constant	-3.55	-3.40	-2.49
		(-2.89)	(-2.86)	(-2.22)
	Long-run
	gdpoecd t-1	3.36	3.34	3.31
		(26.71)	(27.07)	(22.16)
	ree t-1	-1.48	-1.49	-1.52
		(-6.16)	(-6.27)	(-5.08)
	D93(I)*ree t-1	1.38	1.42	1.33
		(3.52)	(3.70)	(2.76)
	ECM	-0.96	-0.95	-0.76
		(-5.33)**	(-5.35)**	(-5.20)**
	Q1	-0.14	-0.14	-0.09
		(-3.00)	(-3.09)	(-2.84)
	Q2	-0.05	-0.05	-0.04
		(-1.83)	(-1.81)	(-1.66)
	Q3	-0.24	-0.24	-0.22
		(-7.66)	(-7.80)	(-7.69)
	R 2 (Aj.)	0.95	0.95	0.95
	R 2	0.96	0.96	0.96
	Notes: See Table 2.

Table 6 . Estimation of the goods imports function with the demand components disaggregated, 1987-1999. FM-PH

 6

	Constant	-7.47	-7.53
		(-10.99)	(-8.56)
	gdpsp		1.02
			(5.67)
	cons	0.97
		(7.46)
	inv	0.12	0.05
		(3.08)	(1.00)
	export	0.70	0.69
		(16.25)	(13.80)
	ree	0.83	0.86
		(5.61)	(5.06)
	ADF-EG	-3.83*	-4.20**
	Notes: See Table 1.
			25

Table 7 . Estimation of the goods imports function with the demand components disaggregated, 1987-1999, 1987-1999. NLLS

 7

	Short-run
	dm t-1	0.03
		(0.18)
	dcons t	0.15
		(0.52)
	dcons t-1	-0.46	-0.57
		(-1.62)	(-2.24)
	dinv t	0.37	0.34
		(5.27)	(5.55)
	dinv t-1	0.12	0.10
		(1.42)	(1.57)
	dexport t	0.42	0.39
		(5.87)	(5.72)
	dexport t-1	0.09
		(0.80)
	dree t	0.54	0.58
		(2.79)	(3.21)
	dree t-1	-0.22
		(-0.97)
	Constant	-3.42	-3.69
		(-2.53)	(-3.69)
	Long-run
	cons t-1	1.74	1.65
		(5.54)	(6.59)
	inv t-1	0.38	0.35
		(4.22)	(5.47)
	export t-1	0.34	0.37
		(3.04)	(4.32)
	ree t-1	-0.15	-0.11
		(-0.51)	(-0.45)
	ECM	-0.57	-0.64
		(-3.11)	(-4.89)**
	Q1	-0.06	-0.04
		(-1.12)	(-1.13)
	Q2	-0.01	-0.01
		(-0.59)	(-0.24)
	Q3	-0.02	0.02
		(-0.40)	(0.37)
	R 2 (Aj.)	0.96	0.96
	R 2	0.97	0.97
	Notes: See Table 2.

Table 8 . Stability parameter test for the imports function with the demand components disaggregated

 8

		Test	P-valor
	SupF	103.39	0.01
	MeanF	21.91	0.01
	Lc	0.43	0.20
	Notes: See Table 3.

Table 9 . Gregory-Hansen test of the C/S model for the goods imports function with the demand components disaggregated

 9 Estimation of the coefficients: y t =α 1 +α 2 ϕ tτ +β 1 cons t +β 2 cons t ϕ tτ +δ 1 inv t +δ 2 inv t ϕ tτ +θ 1 export t +θ 2 export t ϕ tτ +γ 1 ree+γ 2 reeϕ tτ +ε t

	α 1 = -7.92	-7.93
	(-9.27)	(-10.86)
	α 2 = 3.30	4.61
	(1.68)	(2.60)
	β 1 = 0.70	0.71
	(4.84)	(6.19)
	β 2 = 0.14
	(0.52)
	δ 1 = 0.16	0.14
	(2.89)	(3.92)
	δ 2 = -0.01
	(-0.94)
	θ 1 = 0.68	0.69
	(10.16)	(17.30)
	θ 2 = 0.05
		(0.54)
	γ 1 = 1.15	1.15
	(4.44)	(5.93)
	γ 2 = -0.80	-0.98
	(-2.08)	(-2.58)
	Gregory-Hansen test:
	ADF	-4.56	τ=93(II)
		(0.50)
	Z t	-6.56*	τ=93(II)
		(0.20)
	Z α	-47.90	τ=93(II)
		(0.20)
	Notes: See Table 4.

Table 10 . Estimation of the goods imports function with the demand components disaggregated, 1987-1999. NLLS

 10

	Page 29 of 39
	Short-run
	dcons t-1	-0.52	-0.51
		(-2.12)	(-2.16)
	dinv t	0.32	0.32
		(5.28)	(5.38)
	dinv t-1	0.11	0.11
		(1.71)	(1.78)
	dexport t	0.38	0.39
		(5.72)	(6.16)
	dree t	0.84	0.86
		(3.81)	(4.28)
	dree t *D93(II)	-0.66	-0.67
		(-1.91)	(-1.98)
	Constant	-3.62	-3.75
		(-3.76)	(-4.68)
	Long-run
	cons t-1	1.68	1.64
		(6.48)	(8.49)
	inv t-1	0.33	0.31
		(5.09)	(7.39)
	export t-1	0.38	0.39
		(4.30)	(9.37)
	ree t-1	-0.06
		(-0.25)
	ECM	-0.60	-0.61
		(-4.71)*	(-5.10)**
	Q1	-0.06	-0.06
		(-1.51)	(-1.61)
	Q2	-0.01	-0.01
		(-0.42)	(-0.43)
	Q3	-0.003	-0.01
		(-0.08)	(-0.11)
	R 2 (Aj.)	0.96	0.96
	R 2	0.97	0.97
	Notes: See Table 2.

Table 11 . Tests of changes in the integration order

 11

		x	gdpoecd	m	gdpsp	ree	cons	inv	export
	Sup 1 α t	-1.76	-2.00	-2.11	-2.93	-2.30	-2.34	2.27	-1.95
	Mean 1 α t	-0.54	-0.81	-1.22	-2.11	-1.13	-1.73	-1.57 -0.02
	TSup 1 α t	97.1	93.2	93.2	92.3	93.2	89.4	89.1	98.3
	Sup 2 α t	-1.49	-2.20	-1.15	-1.31	-4.32** -1.50	-1.44 -2.18
	Mean 2 α t	-0.76	-0.46	-0.38	-0.15	-1.25	-0.60	-0.65 -0.90
	TSup 2 α t	93.4	93.1	97.4	87.3	92.3	87.3	87.4	94.3
	Sup 1 γ t	-1.78	-2.09	-2.13	-2.93	-2.56	-2.36	-2.44 -2.02
	Mean 1 γ t	-0.67	-0.83	-1.28	-2.22	-1.21	-1.74	-1.75 -0.17
	TSup 1 γ t	90.1	93.1	93.4	94.3	93.1	97.2	97.4	90.4
	Sup 2 γ t	-1.58	-2.28	-1.22	-1.03	-4.45** -1.27	-1.52 -2.20
	Mean 2 γ t	-0.87	-0.44	-0.37	-0.20	-1.37	-0.56	-0.99 -0.99
	TSup 2 γ t	93.4	93.1	97.4	97.4	92.3	89.2	97.1	94.3

Notes: The critical values are tabulated in Fernández and Peruga (1999) *significant at the 5% level; ** significant at the 1% level.

Table 12 . Estimation of the goods exports function with change in the integration order of exchange rate, 1987-1999. NLLS

 12

	Short-run
	dx t-1	0.003
		(0.02)
	dgdpoecd t	0.69
		(0.92)
	dgdpoecd t-1	-1.32	-1.37	-1.48
		(-1.72)	(-1.84)	(-2.01)
	(1-D93(I))*dree t	-0.49	-0.39
		(-1.17)	(-0.99)
	D93(I)*ree t	-0.61	-0.60	-0.56
		(-3.06)	(-3.17)	(-3.03)
	Constant	-2.98	-2.78	-3.28
		(-2.30)	(-2.31)	(-3.01)
	Long-run
	gdpoecd t-1	3.08	3.04	3.07
		(19.06)	(20.04)	(22.16)
	(1-D93(I))*ree t-1	-0.97	-0.98	-0.88
		(-3.52)	(-3.53)	(-3.60)
	ECM	-0.62	-0.61	-0.63
		(-4.34)**	(-4.72)**	(-4.97)**
	Q1	-0.12	-0.12	-0.12
		(-2.40)	(-4.23)	(-4.20)
	Q2	-0.07	-0.07	-0.07
		(-2.81)	(-2.91)	(-2.87)
	Q3	-0.25	-0.25	-0.25
		(-7.68)	(-9.35)	(-9.30)
	R 2 (Aj.)	0.94	0.94	0.94
	R 2	0.95	0.95	0.95
	Notes: See Table 2.

Table 13 . Estimation of the goods imports function with change in the integration order of exchange rate, 1987-1999. NLLS

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

	Page 31 of 39		Submitted Manuscript
	Short-run
	dm t-1	0.002
		(0.01)
	dcons t	0.02
		(0.06)
	dcons t-1	-0.33
		(-1.13)
	dinv t	0.33	0.29	0.29
		(4.70)	(4.29)	(4.41)
	dinv t-1	0.14
		(1.46)
	dexport t	0.43	0.45	0.44
		(5.98)	(6.12)	(6.26)
	dexport t-1			0.08
				(0.67)
	(1-D93(II))*dree t	0.73	0.77	0.76
		(2.54)	(2.86)	(2.94)
	D93(II)*ree t	-0.08	0.04
		(-0.41)	(0.24)
	Constant	-3.05	-2.67	-2.48
		(-1.95)	(-2.45)	(-3.51)
	Long-run
	cons t-1	1.69	1.28	1.33
		(4.59)	(4.44)	(5.90)
	inv t-1	0.31	0.33	0.34
		(3.13)	(3.87)	(4.11)
	export t-1	0.42	0.50	0.49
		(3.82)	(5.06)	(5.71)
	(1-D93(II))*ree t-1	0.14	0.09	0.004
		(0.36)	(0.26)	(0.43)
	ECM	-0.53	-0.48	-0.47
		(-2.60)	(-3.56)	(-4.20)
	Q1	-0.09	-0.02	-0.02
		(-1.49)	(-0.56)	(-0.54)
	Q2	-0.02	0.02	0.02
		(-0.65)	(1.17)	(1.24)
	Q3	-0.01	0.01	0.01
		(-0.22)	(0.27)	(0.30)
	R 2 (Aj.)	0.95	0.95	0.95
	R 2	0.97	0.96	0.96
	Notes: See Table 2.

The order of importance of the three components is the expected one and in consonance with other works. Consumption expenditure, apart from the weight of this type of goods in the total imports (24% in 2000), has the strongest participation in the GDP and its series faithfully follows the evolution of the global aggregated GDP. Investment, as well as being a smaller and more divergent part of income, is intimately related with the purchase of machinery and equipment(19% in 2000). Finally, exports, in spite of also having a reduced participation in the GDP, are linked to the purchase of intermediate products, most of them (57% in 2000), used as inputs in the production of goods that are later exported.

Matsubayashi and Hamori (2003) find evidence of cointegration with structural change for French and German imports with break points in the early nineties, for France at the beginning of 1993 and for Germany at the end of 1991. Nevertheless, there is no cointegration with structural change in the imports of US, Canada, Japan or UK.

This would indicate that the change in the behaviour of prices should only be introduced in the shortterm dynamic, where they are really significant.

17 Even higher than 3 if we were to include the aggregated GDP, a higher value than those derived from the other studies, except for that of[START_REF] García | Funciones trimestrales de exportación e importación para la economía española[END_REF] who reached a coefficient of up to 3.

6.18 The analysis of the change in the integration order of the index of the real effective exchange rate, if it is extended until the second quarter of 2003, giving the same break point[1993(I)]. Moreover, these results are confirmed using consumption prices, a measure of competitiveness less partial and so more unfavourable for the Spanish economy than production

prices.19 [START_REF] Mañez | The decision to export: a panel data analysis for Spanish manufacturing[END_REF] and[START_REF] Moreno | The determinants of Spanish industrial exports to the European Union[END_REF] find a significant relationship between some non-price factors and the decision to export. As well as their direct effect, an effect on price elasticity can be seen in which the bigger the product differentiation the smaller the response of exports to variations in relative price.[START_REF] Alonso | Growth and the external constraint: lessons from the Spanish case[END_REF] notes the relevance of introducing a technological variable that expreses the increase of competitiveness from improvements in quality and, as[START_REF] Balaguer | Structural change in exports and economic growth: cointegration and causality analysis for Spain (1961-2000[END_REF], the structural change in the composition of exports.

research has received the financial support of a CICYT (SEC 2003-06106) Project and the SEIM research group (SEC 269

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