TESTING INTERNATIONAL PRICE TRANSMISSION UNDER ...

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Empirical Analysis: Cointegration Models, Structural Breaks and Policy Reform With the additional outlier model, both for the French and the intervention price it is not possible to reject the null hypothesis of a unit root with structural break, occurred in August 1993. If two breaks are allowed, the situation is much less clear-cut; the 2001:08 break in the intervention price series coincides with the full implementation of Agenda 2000, but doesn’t correspond to the one in the French price. Provided that the change caused by the MacSharry reform is very likely to have had a strong immediate effect rather than a gradual one, as confirmed by visual inspection of the series, for these two prices the innovational outlier model, both allowing for one and two breaks, (according to which the French and the intervention price are both I(0) processes) seems less appropriate, (and indeed, the time of the break moves backwards to a date which has no political sense 80 ). For the US price it is possible to reject the unit root null hypothesis only in the AO model allowing for one break, only. This time the interpretation of the timing of the structural breaks is not immediate. The Zivot and Andrews test has been performed, as well (table 6.3). Both the French and the intervention prices turn out to be I(0) around a broken intercept. Nevertheless, remembering that the null hypothesis of unit root in this test excludes the presence of a structural break, and on the basis of the results of standard unit root tests carried out in the two subsamples identified as before and after the structural break constituted by the MacSharry reform (which, as recalled before, find evidence of nonstationarity; annex B), we conclude that the use of this test is not appropriate for the French and intervention price. For the US price it is not possible to reject the null hypothesis of a unit root process without structural break. Table 6.3 Zivot and Andrews (1992) tests AIC lagmethod TTest lag method Lags Time of t stat 5% Lags Time of t stat 5% the break critical value the break critical value swfrt 1 1993:07 -5.702** -4.80 1 1993:07 -5.702** -4.80 pint t 1 1993:06 -6.042** -4.80 0 1993:06 -5.795** -4.80 hrw t 1 1986:01 -4.686 -4.80 1 1986:01 -4.686 -4.80 AIC/TTest lag method: AIC criterion/5% significance of the t-test used for lag selection ***indicates rejection of the null hypothesis at 1%; ** at 5%; * at 10% significance In order to test for the presence of a unit root in the French price differentiated series, provided the structural break identified in the French price evidently 80 While, for the French price, this might suggest the hypothesis of some “rational expectations” behaviour of market operators whose actions anticipate the following reduction in the intervention price, this is actually not the case since the same occurs for the intervention price, which is the very policy variable. 99
Empirical Analysis: Cointegration Models, Structural Breaks and Policy Reform corresponds to the MacSharry reform, reference can be made to standard unit root tests which have been repeated on the log-differentiated series in the two subsamples identified as before and after the MacSharry reform (see annex B). For the US price, which most plausibly seems to be a I(1) process with no relevant structural breaks, evidence of its first-difference stationarity is confirmed by the conventional unit root tests reported in annex B. 6.4 A cointegration model allowing for structural breaks At this point, the Johansen, Mosconi and Nielsen (2000) model has been tested on the data 81 . The model presented in paragraph 3.3.2.3, that is 100 ⎡β⎤ ' ⎡ pt −1 ⎤ k−1 k q ∆pt = α⎢ + γEt + Γ ∆pt i k D j, t k i ε i 1 1 + i 1 j 2 i, j + t µ ⎥ ⎢ tE ⎥ ∑ = − ∑ = ∑ = + − ⎣ ⎦ ⎣ t−1 ⎦ (6.1) has been estimated restricting the broken level to the cointegration space 82 , the only difference being the insertion of a constant term and the corresponding elimination of one of the q dummy variables used. This means that the coefficients of the dummies indicating the policy regimes will have to be interpreted as relative to the constant valid in the overall period. Throughout this chapter, the underlying assumption is that the rank of the cointegration matrix between the French and the US price series, once the various structural breaks have been taken into account, is one. And this because in the estimated models more than 2 structural breaks will be introduced, differently from the Johansen, Mosconi and Nielsen (2000) procedure, which doesn’t allow to test for the cointegration rank with more than 2 structural breaks. According to conventional Johansen’s test, the EU and the US price are not cointegrated (when tested in the restricted constant option with 5 lags 83 , the null hypothesis of a zero rank in the cointegration matrix yield a trace and λ-max statistics of 16.086, p-value 0.173, and 12.538, p value 0.161, respectively). Nonetheless, standard ADF tests have been repeated on all the long-run residuals of the models estimated and confirm that they are I(0) at 5% level of significance. 81 The prices have been de-seasonalized with the TRAMO/SEATS method developed by V. Gomez, A. Maravall (http://www.bde.es/servicio/software/econome.htm), in order to remove autocorrelation in the residuals. Unit root tests have been repeated on the de-seasonalized series and are reported in annex E. 82 I.e., t = γ = 0. Indeed, the series don’t show any linear trend in levels, but the presence of a constant in the cointegration space is needed to account for all those factors contributing to price differentials not explicitly modelled. 83 The optimum number of lags selected for the underlying VAR is 5 according to the SBIC.
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TESTING INTERNATIONAL PRICE TRANSMI
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Associazione “Alessandro Bartola
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Index 1 INTRODUCTION ..............
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1 INTRODUCTION The notion of “pri
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Introduction These recent phenomena
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Introduction Whereas the Common Agr
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Price Transmission and the Law of O
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3 EMPIRICAL TESTS FOR SPATIAL PRICE
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Empirical Tests for Spatial Price A
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= β 2 i0 Empirical Tests for Spati
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∆p + t ∑ ⎡β⎤ ' ⎡ p t =
Page 50 and 51: Empirical Tests for Spatial Price A
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Page 54 and 55: 4 INTERNATIONAL SOFT WHEAT MARKETS
Page 56 and 57: International Soft Wheat Markets Un
Page 58 and 59: 30.000 20.000 10.000 0 1961 1963 19
Page 70 and 71: 250 200 150 100 50 1978:12 Internat
Page 76 and 77: 5 EMPIRICAL ANALYSIS: COINTEGRATION
Page 78 and 79: Empirical Analysis: Cointegration M
Page 82 and 83: α LM test ARCH(13) Empirical Analy
Page 96 and 97: 6 EMPIRICAL ANALYSIS: COINTEGRATION
Page 110: Empirical Analysis: Cointegration M
Page 113 and 114: Concluding Remarks These models rep
Page 116 and 117: REFERENCES Alderman, H., 1992. Inte
Page 118 and 119: References agricultural policies: s
Page 120 and 121: References Isard, P., 1977. How far
Page 122: References Thompson, S. R., Sul, D.
Page 125 and 126: Annexes 124 Annex A - Empirical ana
Page 127 and 128: Annexes METHODOLOGY DATA USED MAJOR
Page 139 and 140: Annexes Variable swfr 138 Annex B -
Page 141 and 142: Annexes Variable swfr 140 Table B.3
Page 143 and 144: Annexes 142 Table B.7 - Unit root t
Page 145 and 146: Annexes 144 Table C.5 - Cointegrati
Page 147 and 148: Annexes 146 Table D.4 - Unit root t
Page 149 and 150: Annexes 148 Annex E - Unit root tes
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Annexes Table E.5 - Unit root tests
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The notion of “price transmission
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