Troels K Bergmann
email: tbergmann@health.sdu.dk

Henrik Gréen

Charlotte Brasch-Andersen

Mansoor R Mirza

Jørn Herrstedt

Berit Hølund

Per Damkier Andreas Bois

Werner Vach

Kim Brosen

) H Gréen

Retrospective study of the

Keywords: paclitaxel, CYP2C8, ABCB1, ovarian cancer, neutropenia, neuropathy

de niveau recherche, publiés ou non, émanant des établissements d'enseignement et de recherche français ou étrangers, des laboratoires publics ou privés.

Introduction

Paclitaxel is a highly active anticancer drug useful in the treatment of patients with e.g. breast, lung or ovarian cancer. It was first isolated in the seventies from the bark of the western yew, Taxus brevifolia [START_REF] Wani | Plant antitumor agents. VI. The isolation and structure of taxol, a novel antileukemic and antitumor agent from Taxus brevifolia[END_REF]. In advanced ovarian cancer the initial response rate is almost 80 % but unfortunately most patients eventually relapse [START_REF] Karlan | Gynecologic Cancers[END_REF]. Patients are allocated to treatment according to the FIGO stage (International Federation of Gynecology and Obstetrics) with post-surgery chemotherapy with a taxane and a platin being the widely accepted treatment for stage II-IV. Paclitaxel is dosed normalized to body surface area (BSA) and in most regimens infused over 1, 3 or 24 hours. Dose limiting toxicities are neutropenia and neuropathy; other toxicities include arthralgia, myalgia and alopecia whereas nausea and vomiting are experienced by less than 30 % of the patients. Toxicity and clinical effects of paclitaxel vary greatly among patients and remain a clinically relevant problem with implications on survival and quality of life. Toxicity can result in dose delay, dose reduction or even early cessation of the treatment. It has been proposed and investigated that single nucleotide polymorphism (SNPs) could explain the variability in toxicity and effect. Sequence variants in the genome might have an impact on clinical outcome either indirectly by altering the elimination and/or disposition or directly by changes in the drug concentration in the micro environment of the cells in the different tissues.

Paclitaxel is metabolized to inactive compounds by CYP2C8 and CYP3A4 in the liver [START_REF] Harris | Metabolism of taxol by human hepatic microsomes and liver slices: participation of cytochrome P450 3A4 and an unknown P450 enzyme[END_REF][START_REF] Rahman | Selective biotransformation of taxol to 6 alpha-hydroxytaxol by human cytochrome P450 2C8[END_REF][START_REF] Kumar | Comparative in vitro cytotoxic effects of taxol and its major human metabolite 6 alpha-hydroxytaxol[END_REF], and is also a substrate of the ATP driven efflux pump P-glycoprotein encoded by the ABCB1 gene [START_REF] Sparreboom | Limited oral bioavailability and active epithelial excretion of paclitaxel (Taxol) caused by Pglycoprotein in the intestine[END_REF]. This study was initiated based on reports on CYP2C8*3 as having decreased paclitaxel metabolism [START_REF] Dai | Polymorphisms in human CYP2C8 decrease metabolism of the anticancer drug paclitaxel and arachidonic acid[END_REF][START_REF] Soyama | Non-synonymous single nucleotide alterations found in the CYP2C8 gene result in reduced in vitro paclitaxel metabolism[END_REF] and reports of a functional significance of ABCB1 SNPs with particular interest in C1236T, G2677T/A and C3435T [START_REF] Tanabe | Expression of P-glycoprotein in human placenta: relation to genetic polymorphism of the multidrug resistance (MDR)-1 gene[END_REF][START_REF] Kim | Identification of functionally variant MDR1 alleles among European Americans and African Americans[END_REF][START_REF] Hoffmeyer | Functional polymorphisms of the human multidrug-resistance gene: multiple sequence variations and correlation of one allele with P-glycoprotein expression and activity in vivo[END_REF]. In the study design these four 'highly implicated' and relatively frequent SNPs were selected (pre hoc) for the primary analysis; in addition other candidate genes linked to the pharmacodynamics and pharmacokinetics of paclitaxel were chosen for an exploratory analysis namely CYP3A5, ABCC1, ABCC2, ABCG2, ABCC10, CYP1B1 and SLCO1B3 [START_REF] Marsh | Taxane pharmacogenetics[END_REF][START_REF] Huisman | MRP2 (ABCC2) transports taxanes and confers paclitaxel resistance and both processes are stimulated by probenecid[END_REF][START_REF] Mcfadyen | Cytochrome P450 CYP1B1 over-expression in primary and metastatic ovarian cancer[END_REF][START_REF] Marsh | Pharmacogenetic Assessment of Toxicity and Outcome After Platinum Plus Taxane Chemotherapy in Ovarian Cancer: The Scottish Randomised Trial in Ovarian Cancer[END_REF][START_REF] Smith | Variants in the SLCO1B3 gene: interethnic distribution and association with paclitaxel pharmacokinetics[END_REF][START_REF] Hopper-Borge | Analysis of the drug resistance profile of multidrug resistance protein 7 (ABCC10): resistance to docetaxel[END_REF][START_REF] Mcfadyen | Cytochrome P450 CYP1B1 protein expression: A novel mechanism of anticancer drug resistance[END_REF]. These are all drug transporters except CYP3A5 which has a substrate range similar to CYP3A4, and CYP1B1 which has been implied to play a role in taxane metabolism and therapeutic effects [START_REF] Marsh | Taxane pharmacogenetics[END_REF].

The aim of the study was primarily to evaluate the impact of the CYP2C8*3, ABCB1 C1236T, G2677T/A and C3435T variants on paclitaxel toxicity and survival and secondarily to explore the role of the other variants in CYP2C8 and ABCB1 as well as the variants in CYP1B1, CYP3A4/5, SLCO1B3, ABCC1, ABCC2, ABCG2 and ABCC10. We present a retrospective pharmacogenetic study of 119 Scandinavian Caucasian patients who were treated with paclitaxel and carboplatin in either the AGO-OVAR-9 trial (ClinicalTrials.gov identifier: NCT00052468) or the NSGO-OC9804 trial (NCT00004934) (phase III trials of first-line treatment of ovarian cancer).

Patients and methods

Patients

In this study clinical data and tissue was collected from 119 patients (Table 1). The patients were all previously treated with paclitaxel (175mg/m 2 ) and carboplatin (AUC 5-6, using Calvert´s formula) in either the AGO-OVAR-9 trial (OVAR-9) or the NSGO-OC9804 trial (TEC). Both trials compared a three compound regimen to the standard treatment of paclitaxel/carboplatin in patients with primary epithelial ovarian cancer. In the OVAR-9 trial the primary and secondary endpoints were overall survival and progression free survival, response rate, duration of response, toxicity and quality of life.

The study closed for inclusion in April 2004. A total of 83 patients in Denmark and Sweden were randomized to paclitaxel and carboplatin. In the TEC trial the primary and secondary endpoints were progression free survival and overall survival, toxicity and quality of life. The study closed for inclusion October 2001. A total of 99 patients in Denmark and Sweden were randomized to paclitaxel and carboplatin. Treatment records and follow-up data for OVAR-9 and TEC are kept by the AGO Ovarian Cancer Study Group in Wiesbaden, Germany and by the NSGO in Odense, Denmark.

Formalin fixed paraffin embedded tissue samples from the patients were collected from departments of pathology at the participating centers in Denmark and Sweden. Eligibility criteria for this study were: 1) participation in either the OVAR-9 or the TEC study in either Denmark or Sweden, 2) randomization to paclitaxel and carboplatin treatment, 3) availability of formalin fixed paraffin embedded tissue (any type available) and 4) availability of treatment records and follow-up data. No patients were contacted. The study was approved by the regional ethics committee of Southern Denmark and the regional ethics committee in Linkoping, Sweden.

Toxicity

Patients were originally evaluated for toxicity using Common Toxicity Criteria (CTC-NCIC). The two studies used revisions from March 1998 and December 1994 which have only insignificant differences regarding the chosen toxicities. Generally, grade 0 was used for absence of the toxicity, 1, 2, 3 and 4 were assigned for mild, moderate, severe toxicities and very severe toxicity. The principal characteristic of grade 3 and 4 is interference with normal daily activity as opposed to grade 1 and 2.

Summary statistics of toxicities were evaluated pre hoc with regard to frequency distribution and cross-wise correlation. Overall survival and two toxicities with no indication of being strongly correlated (Spearman correlation coefficient = 0.07) were picked for primary analysis: 1) Sensoric neuropathy: grades seemed to be increasing with increasing number of cycles, and so time-to-event analysis was used, with time being the cycle number (0 to 9) and event time being the first cycle with reported increase in sensoric neuropathy score (CTC) of two from baseline, 2) Neutrophil depression calculated as difference between baseline and nadir divided by the baseline value in %. Nadir was defined as the lowest observed value between two cycles regardless of the actual sampling time.

Seven other toxicities of which none seemed to be strongly cross-wise correlated (Spearman coefficient ≤ 0.40) were left for explorative analysis: myalgia, arthralgia, mucositis, hearing loss, vomiting/nausea, motor neuropathy and compliance. For all toxicities the highest grade observed for any cycle was used. Nausea and vomiting seemed to be correlated (Spearman coefficient of 0.66) and the highest grade for either of the two was used. The grading for compliance was constructed using the actual dose of paclitaxel administered. Grade 1 was defined as at least one episode of reduced paclitaxel dose (range of 50 % -90 % of initial dose) and grade 2 as occurrence of at least one cycle where paclitaxel was not given. In two patients who experienced an allergic reaction during paclitaxel infusion only the subsequent cycles were evaluated for compliance. Evaluation of all toxicity data were completed and inserted in the database, which was locked, before the analyses of genetic variants were disclosed. Summary of the toxicity is presented in Table 2.

Genotyping procedures

Genomic DNA was extracted from formalin fixed paraffin embedded tissue using QIAamp DNA Mini Kit, tissue protocol (VWR International A/S, Denmark) according the manufacturers protocol.

SNPs in CYP2C8, CYP3A4, SLCO1B3 (except G767C) and ABCB1(except A-1G) were determined using Pyrosequencing as previously described [START_REF] Green | mdr-1 single nucleotide polymorphisms in ovarian cancer tissue: G2677T/A correlates with response to paclitaxel chemotherapy[END_REF][START_REF] Green | Pharmacogenetic studies of Paclitaxel in the treatment of ovarian cancer[END_REF][START_REF] Green | ABCB1 G1199A polymorphism and ovarian cancer response to paclitaxel[END_REF]. The variants in CYP1B1, CYP3A5, ABCC1, ABCC10 and G767C in SLCO1B3, A-1G in ABCB1 and rs2273697 in ABCC2 were genotyped using TaqMan® pre-designed SNP genotyping assays by Applied Biosystems (Foster city, CA). SNPs in ABCG2 and rs17222723 and rs8187710 in ABCC2 were genotyped using SYBRgreen real time PCR assays as previously described [START_REF] Bergmann | Impact of CYP2C8*3 on paclitaxel clearance: a population pharmacokinetic and pharmacogenomic study in 93 patients with ovarian cancer[END_REF].

Statistics

In order to overcome the problem of multiple testing we split the analysis into a primary and an exploratory analysis. For the primary analysis two sided P-values with a significance level of 0.05 was used. P-values were calculated for the exploratory analysis in similar fashion and reported without any correction for multiple testing -the results should therefore be interpreted with caution. Briefly, we assumed that heterozygocity is an intermediate between the homozygotes and that CTC scores are on an ordinal scale we thus used a non parametric test for trend (STATA10 command: nptrend, 2009, StataCorp, Texas, USA) for SNPs with three genotypes and similarly the Wilcoxon rank sum test for those with two genotypes. For the tri-allelic SNP G2677T/A in ABCB1 patients carrying an A allele was omitted from the analysis. The Kaplan-Meier method with Log-rank test was used to test the risk of developing sensoric neuropathy (STATA10 command: sts test, with trend option). A Spearman correlation coefficient > 0.5 was arbitrarily set to determine if different toxicities were correlated.

Overall survival analysis was carried out fitting a Cox model under the assumption of proportional hazard. This assumption was tested visually and by Schoenfeld residuals (data not shown). Classical prognostic variables [START_REF] Karlan | Gynecologic Cancers[END_REF] were tested one at a time in a univariate analysis (Table 3). Age ( 65 or >65 years), CA125 levels at baseline ( 35 or >35 U/ml) and FIGO stage (trend from stage II to IV) were picked for the final model in which the SNPs were tested for trend. All calculations were carried out in STATA 10 (2009, StataCorp, Texas, USA).

Results

Patients

One hundred nineteen of 182 patients were eligible for analysis; 56 patients from the OVAR-9 trial and 63 from the TEC trial. Twenty and 31 patients were ineligible from the OVAR-9 and TEC trial respectively, because of oncology departments not participating. Tissue was unavailable from a total of 12 patients from the participating departments. Characteristics are summarized in Table 1.

Results of genotyping

Twenty-two single nucleotide polymorphisms (SNPs) were analyzed in 10 genes. For 57 patients both tumor and non cancerous tissue was available. The total call rate (across all SNPs) was 94.7 % evenly distributed (data not shown). Mismatching genotypes were found evenly distributed in 2.3 % of the cases where both tumor and non cancerous tissue was available. In these cases no final genotype was called. In one patient all calls from the tumor sample were discarded because the DNA quality was obviously very poor with a call rate of 41 % across the 22 SNPs. Allele frequencies were all similar to frequencies reported for Caucasians on the NCBI SNP database (dbSNP). The genotype distributions were all in Hardy-Weinberg equilibrium except CYP2C8*4, *1B and *1C, and ABCB1 A61G.

Toxicity

Toxicity registration was available from 710 treatment cycles (Table 2). Twenty-three patients received treatment beyond the 6 th cycle. One patient did not receive chemotherapy (no cause stated).

Toxicity registration for this patient was not done, but survival data was included in the final analysis (intention to treat).

Primary analysis

No statistical significant correlations was found between the SNPs: CYP2C8*3, ABCB1 C1236T, G2677T/A and C3435T and neutrophil depression and sensoric neuropathy (Table 4).

For overall survival the calculated hazard ratio for the individual genotype as a covariate in the Cox model is reported along with the 95% confidence interval (Table 4). None of the SNPs selected for primary analysis were statistically significantly associated with overall survival of the patients. ABCB1 G2677T/A which is a tri-allelic SNP was tested excluding patients carrying an A allele. For completeness analyses treating the A allele as a T allele and using the five genotypes as a categorical variable were also carried out.

Exploratory analysis

Among the 190 tests of association between particular toxicities and SNPs, seven were found to have a P-value ≤ 0.05 (data not shown). ABCB1 G1199A was associated with increased risk of death with a hazard ratio of 2.03 (95% ci: 1.08-3.83), P-value = 0.03 (Figure 1).

Discussion

Definite results linking frequent ABCB1 variants to taxane toxicity and effects in vivo have been elusive. This study retrospectively investigates possible correlations between a wide selection of candidate genes all with a reputed role in either the transport or the metabolism of paclitaxel and the toxicity and survival of patients treated with paclitaxel. Clinical data and formalin fixed paraffin embedded tissue was gathered from 119 patients with ovarian cancer who were treated with paclitaxel and carboplatin in either the OVAR-9 or the TEC trial. The genetic variant CYP2C8*3 (a paclitaxel metabolizing enzyme) and three variants in the drug transporter ABCB1: C1236T, G2677T/A and C3435T were picked pre hoc for primary analysis; no statistical significant correlation for sensoric neuropathy, neutrophil depression and overall survival was found. This finding contrasts the correlation reported by Nakajima et al [START_REF] Nakajima | Pharmacokinetics of paclitaxel in ovarian cancer patients and genetic polymorphisms of CYP2C8, CYP3A4, and MDR1[END_REF] for leukocytopenia and C3435T in 23 Japanese women with ovarian cancer and the report of Sissung et al [START_REF] Sissung | Association of ABCB1 genotypes with paclitaxel-mediated peripheral neuropathy and neutropenia[END_REF] who found a trend between neutrophil decrease and the ABCB1 C3435T/G2677T/A compound genotype in 26 patients with advanced solid tumors. Interestingly Green et al [START_REF] Green | Pharmacogenetic studies of Paclitaxel in the treatment of ovarian cancer[END_REF] recently reported a correlation between CYP2C8*3 and hematological toxicity.

We also conducted a wide exploratory analysis of 19 SNPs in a total of 10 genes and both sensoric neuropathy, neutrophil depression and overall survival and other toxicities that are related to paclitaxel treatment i.e. myalgia, arthralgia, mucositis, hearing loss, motor neuropathy, vomiting/nausea and compliance. This analysis encompasses some 190 tests (data not shown) of which seven had a P-value ≤ 0.05. This finding is likely due to chance alone, given the multiple testing. This multitude of statistical testing can serve a purpose for future investigations but have no value without being confirmed in other studies. None of the seven possible associations have to our knowledge been reported in the literature before. In the explorative analysis of overall survival, an association with the non synonymous G1199A SNP in the ABCB1 transporter was found with a Pvalue of 0.03 (uncorrected) and a hazard ratio of 2.03 [95 % c.i. 1.08-3.83] favoring patients with 1199GG genotype (Figure 1). This is in accordance with the shorter progression free survival for patients carrying the 1199GA genotype found by Green et al [START_REF] Green | ABCB1 G1199A polymorphism and ovarian cancer response to paclitaxel[END_REF]. Similarly Johnatty et al [START_REF] Johnatty | ABCB1 (MDR 1) Polymorphisms and Progression-Free Survival among Women with Ovarian Cancer following Paclitaxel/Carboplatin Chemotherapy[END_REF] found a better progression-free survival for patients carrying the 2677TA alleles and Gréen et al [START_REF] Green | mdr-1 single nucleotide polymorphisms in ovarian cancer tissue: G2677T/A correlates with response to paclitaxel chemotherapy[END_REF] found an association for the same variant and good response to paclitaxel treatment. The finding should be confirmed in other studies before any conclusions can be made. CYP1B1*3 has previously been associated with survival after paclitaxel treatment in breast cancer [START_REF] Gehrmann | Prediction of paclitaxel resistance in breast cancer: is CYP1B1*3 a new factor of influence?[END_REF][START_REF] Marsh | Pharmacogenetic analysis of paclitaxel in breast cancer[END_REF]; but this finding was not reproduced in our material of ovarian cancer.

In 2007 Marsh et al [START_REF] Marsh | Pharmacogenetic Assessment of Toxicity and Outcome After Platinum Plus Taxane Chemotherapy in Ovarian Cancer: The Scottish Randomised Trial in Ovarian Cancer[END_REF] reported a very comprehensive retrospective pharmacogenetic study of toxicity and response to treatment with a taxane in 914 ovarian cancer patients. No significant association was found for a selection of variants in the candidate genes: ABCB1, ABCC1, ABCG2, CYP1B1, CYP2C8 and CYP3A4/5 and hematologic toxicity, neurotoxicity and gastrointestinal toxicity. While this study included far more patients than our study and other previous studies it also raised a debate concerning power and the interpretation of P-values [START_REF] Vach | No evidence for taxane/platinum pharmacogenetic markers: just lack of power?[END_REF][START_REF] Branford | Ethnic considerations in pharmacogenetic studies[END_REF][START_REF] Maitland | Interpreting P values in pharmacogenetic studies: a call for process and perspective[END_REF]. Our results however, do agree with the conclusions made by Marsh et al; but are again contrasted by a very recent paper by Leskelä et al [START_REF] Leskela | Polymorphisms in cytochromes P450 2C8 and 3A5 are associated with paclitaxel neurotoxicity[END_REF] who demonstrates an association between CYP2C8*3 and CYP3A5 in a mixed retrospective/prospective study.

In this study we investigate candidate genes with putative impact on the pharmacokinetics of paclitaxel under the assumption that these might influence on toxicity or survival either directly through impact on clearance or the distribution of paclitaxel or indirectly through the paclitaxel concentration in the micro environment in the tumor and susceptible tissues (bone marrow, peripheral nerves etc.). We have thus not addressed the possibility that the susceptibility of the mentioned targets by themselves is determined by genetic variation. Ideally, future pharmacogenetic studies of chemotherapy should try to combine the pharmacokinetic and the pharmacodynamic components.

In conclusion this study primarily investigated the notion that the genetic variants CYP2C8*3 and ABCB1 C1236T, G2677T/A and C3435T explain some of the observed variability in toxicity and survival of patients treated with paclitaxel and carboplatin in ovarian cancer. No statistical significant associations were found in the primary analysis. The study also involved an explorative analysis of other candidate genes which might serve a purpose for studies in the future. a) Genotyping data missing for some patients b) One patient did not receive chemotherapy and was excluded from the toxicity analysis but included in the survival analysis (intention to treat) c) Two patients with the A allele were excluded from the analysis d) Hazard ratios and 95% confidence intervals for each SNP added to the final Cox model with FIGO stage, age and CA125 as covariates

a)

  One patient did not receive chemotherapy b) No paclitaxel reduction c) Paclitaxel dose reduced to <90% of initial dose d) Paclitaxel discontinued e) One patient had grade 4 nausea/vomiting f) Na = not defined

Figure 1 -

 1 Figure 1 -Overall survival function a (fraction) for 119 patients as function of ABCB1 G1199A genotype. Hazard ratio 2.03 (95%c.i 1.08;3.83), P-value 0.03.

Table 1 -

 1 Summary of the 119 patients in the study

	N=119 evaluable patients	Value -median (range)
	Age (years)	57.8 (26.1-77.4)
	BSA a (m 2 )	1.71 (1.4 -2.24)
		Value -N (%)
	WHO/ECOG b Performance status
	0	64 (53.8 %)
	1	52 (43.7 %)
	2	3 (2.5 %)
	Tumor type
	Ovary	113 (95 %)
	Peritoneal	2 (1.7 %)
	Fallopian tube	4 (3.4 %)
	FIGO c stage
	IIb	10 (8.4 %)
	IIc	8 (6.7 %)
	IIIa	8 (6.7 %)
	IIIb	8 (6.7 %)
	IIIc	63 (52.9 %)
	IV	22 (18.5 %)
	Tumor grade (G)
	Well differentiated (G1)	13 (10.9 %)
	Moderately differentiated (G2)	30 (25.2 %)
	Poorly or undifferentiated (G3)	55 (46.2 %)
	Undetermined	21 (17.7 %)
	a) Body surface area (unavailable for 1 patient).
	b) Eastern Cooperative Oncology Group
	c) International Federation of Gynecology and Obstetrics

Table 2 -

 2 Common Toxicity Criteria (CTC) grades for N=119 a patients

	CTC grade	0		1		2		3
		N		N		N		N
	Sensoric neuropathy	37	31 %	58	49 %	20	17%	3	3 %
	Arthralgia	55	47 %	37	31 %	24	20%	2	2 %
	Myalgia	52	44 %	34	29 %	29	25 %	3	3 %
	Febrile neutropenia	117	99 %	na f		na f		1	1 %
	Mucositis	87	74 %	27	23 %	4	3 %	na f
	Hearing loss	106	90 %	10	8 %	2	2 %	na f
	Nausea/vomiting	41	35 %	53	45 %	20	17 %	4 e	3 %
	Neuropathy (ataxia)	83	70 %	25	21 %	9	8 %	1	1 %
	Compliance b,c,d	99 b	84 %	14 c	12 %	5 d	4 %	na f

Table 3 -

 3 Cox proportional hazard model for overall survival for N=119 a patients. FIGO stage, age and Cancer antigen 125 were included as covariates in the final model.

	Covariate		Patients	Univariate analysis	Multivariate analysis
										P-
				N	%	HR	b	P-value	HR	b	95%ci	value
	FIGO	c stage		trend		2.3	<0.0001	1.92 1.26-2.93 0.002
				by category
				II 18 15	1
				III 79 66	2.28	0.032
				IV 22 18	5.29	<0.001
	Histology		Serous 78 66	1				na
				Mucinous 8	7	3.95	<0.001
				Clear cell 2	2	3.36	0.096
				Endometroid 14 12	0.72	0.426
			Undifferentiated 2	2	0.66	0.676
			Undetermined 15 13	1.76	0.081
	Tumor grade	d	G1 13 13	1				na
				G2 30 31	2.03	0.158
				G3 55 56	2.41	0.065
	Residual tumor	<1 cm 50 42	1.96	0.005		na
				1cm 69 58
	Age (years)		65 83 70	2.49	0.0003	2.36 1.47-3.80 <0.001
				>65 36 30
	Performance status	0 64 54	1.35	0.18		na
				1-2 55 46
	Cancer antigen 125	35 U/ml 24 20	2.52	0.005	1.96 1.01-3.83 0.05
				>35 U/ml 95 80
	a) One patient did not receive treatment but were not excluded from the survival analysis
		(intention to treat)
	b) Hazard ratio
	c) International Federation of Gynecology and Obstetrics
	d) Patients with missing grade were excluded from the analysis

Table 4 -

 4 Association tests (non parametric trend tests) of toxicity and overall survival versus SNPs in 119 a patients.

								Toxicity b and overall survival
		Gene variant			Genotype a
							P-values and hazard ratio d with 95% confidence interval
	Gene	dbSNP ID	Variant	Wildtype	Heterotype	Variant	Neutrophil decrease%	Neuropathy logRank test	Overall survival Hazard ratio (95% ci)
	CYP2C8	rs10509681	A1196G(*3)	95	21	1	0.64	0.98	1.13 (0.66-1.92)
	ABCB1	rs1128503	C1236T	31	55	26	0.66	0.10	1.18 (0.84-1.66)
		rs2032582	G2677T/A c	35	49(GT) 1(GA)	28 (TT) 1(AA)	0.77	0.51	1.16 (0.84-3.5)
		rs1045642	C3435T	27	48	39	0.54	0.48	0.95 (0.71-1.26)

Acknowledgements

We wish to acknowledge the AGO Ovarian Cancer study group in Wiesbaden, Germany and The Nordic Society of Gynecologic Oncology, Clinical Trial Unit, Odense, Denmark for making available the clinical data from the AGO-OVAR-9 study and the NSGO-OC9804 (TEC) study.

We wish to acknowledge the work and invaluable help of lab technicians Ingrid Jakobsen Falk and Karin Skoglund in Linköping and Pernille Jordan in Odense. The work was financially supported by grants from European Commission (CHEMORES LSHC-CT-2007-037665), the Swedish Cancer Society, the Swedish Medical Society-Linköping branch, the County Council in Östergötland, the Danish Ministry of Interior Affairs and Health (2001-2007)(J.nr 2006-12103-276), the Danish Research Agency (J.nr 271-05-0266), Borgholm Rotary club, Sweden and Roche Denmark, Hvidovre, Denmark.

Conflicts of Interest

The authors declare that they have no conflict of interest.


Rosa M Antonijoan

Ignasi Gich

Analia Azaro

Sergio Sainz

Joaquim Balanzó

Iñaki Izquierdo

Javier Borja
email: javier.borja@uriach.com

Esther Donado

Iris Blanch

Manel J Barbanoj

J Uriach

Gastrointestinal safety of triflusal solution in healthy volunteers: a proof of concept endoscopic study

Keywords: Triflusal, antiplatelet agents, gastrointestinal safety, clinical trials

de niveau recherche, publiés ou non, émanant des établissements d'enseignement et de recherche français ou étrangers, des laboratoires publics ou privés.

Introduction

Triflusal is an antiplatelet agent marketed in 1981 as capsules containing 300 mg of active substance. Triflusal irreversibly acetylates cyclooxygenase isoform 1 (COX-1) and therefore inhibits thromboxane biosyntesis [1,[START_REF] Murdoch | Triflusal: a review of its use in cerebral infarction and myocardial infarction, and as thromboprohylaxis in atrial fibrillation[END_REF] . Unlike acetylsalicylic acid (ASA), triflusal inhibits phosphodiesterase, the enzyme responsible for degrading cAMP and cGMP, both of which have antiaggregant effects [START_REF] García-Rafanell | Effect of triflusal and other salicylic acid derivatives on cyclic AMP levels in rat platelets[END_REF]. Also unlike ASA, the endothelial synthesis of prostacyclin is preserved with triflusal [1]. The authorized posology of triflusal capsules is 600 mg (two capsules) in single daily dose or 900 mg (three capsules) in fractionated daily doses.

In controlled clinical trials in patients with acute myocardial infarction [START_REF] López-Bescos | Randomized comparative trial of Triflusal and aspirin following acute myocardial infarction[END_REF] and in patients with ischemic stroke or transient ischemic attack [START_REF] Matias-Guiu | Comparison of Triflusal and aspirin for prevention of vascular events in patients after cerebral infarction: the TACIP Study: a randomized, double-blind, multicenter trial[END_REF][START_REF] Culebras | Triflusal vs aspirin for prevention of cerebral infarction[END_REF] triflusal administered at a dose of 600 mg daily has demonstrated an efficacy similar to ASA in the prevention of vascular events but with a lower hemorrhagic risk. These results were confirmed in a metaanalysis [START_REF] Costa | Triflusal for preventing serious vascular events in people at high risk[END_REF] . Also, the efficacy and tolerability of triflusal in patients with unstable angina [START_REF] Plaza | Protective effect of triflusal against acute myocardial infarction in patients with instable angina: results of a Spanish multicenter trial[END_REF], aortocoronary bypass [9], bioprosthetic valve replacement [10] and peripheral arterial disease [11], has been evidenced in controlled clinical trials. The association of triflusal and moderate intensity oral anticoagulation is more effective than standard oral anticoagulation to prevent vascular events in patients with atrial fibrillation [12] . Triflusal is well tolerated in patients with ASA-induced asthma [13].

Gastrointestinal tolerability of triflusal is good. In the Triflusal versus Aspirin in Cerebral Infarction Prevention (TACIP) study [START_REF] Matias-Guiu | Comparison of Triflusal and aspirin for prevention of vascular events in patients after cerebral infarction: the TACIP Study: a randomized, double-blind, multicenter trial[END_REF] , performed in 2113 patients that suffered an ischemic stroke or transient ischemic attack in the previous six months, the incidence of gastrointestinal hemorrhage was lower with triflusal than with aspirin ( 4.6% vs 7.6%; p = 0.005) as it was the incidence of peptic ulcer (0.1% vs 0.8%; p = 0.021). However, the incidence of dyspepsia was higher with triflusal than with aspirin (27.4% vs 21.7%; p = 0.002). The previously mentioned metaanalysis [START_REF] Costa | Triflusal for preventing serious vascular events in people at high risk[END_REF] evidenced that gastrointestinal hemorrhage was higher with aspirin than with triflusal (OR: 1.83; 95% CI: 1.35 to 2.48) whereas nonhemorrhagic gastrointestinal adverse events were lower with aspirin than with triflusal (OR: 0.84; 95% CI. 0.75 to 0.95) 7 . In case-control studies [14,15] triflusal has not been associated with a significant risk of gastrointestinal hemorrhage.

The primary aim of this study is to compare the gastrointestinal safety of triflusal oral solution with triflusal capsules in healthy volunteers by means of upper endoscopy evaluation. As secondary objectives the gastrointestinal safety of triflusal oral solution and triflusal capsules were compared with placebo in a wide sample of healthy volunteers.

Subjects and Methods

Subjects

Sixty healthy subjects (31 men and 29 women; mean age, 25.8 years; range, 20-42 years) participated in the study. Before entering the study, all subjects were required to provide a complete medical history and undergo a physical examination, electrocardiogram (ECG) and clinical laboratory tests. Subjects were excluded from the study if they met some of the following criteria: history of gastrointestinal disease or current gastrointestinal symptoms; abnormal results concerning vital signs (systolic and diastolic blood pressure, heart rate and body temperature) or ECG; being smoker; alcohol or drug abuse in the previous month or positive urine test for ethanol, cannabis, cocaine, amphetamines, benzodiacepines or opiates; antecedent of hypersensitivity to drugs; consumption of stimulant beverages (more of five coffees, teas or cola beverages at day) in the previous week to the endoscopic examination; intake of any other medication (including over the counter and herbal medicines) in the two previous weeks; seropositivity for hepatitis B or C viruses or HIV; positive test for Helicobacter pylori; history of cardiovascular, respiratory, renal, hepatic, endocrine, hematologic or neurological disease or other chronic diseases, surgical intervention in the previous six months, pregnancy or women not using an effective contraceptive method, participation in other clinical trial during the previous three months, being blood donors in the previous four weeks. Moreover, subjects were excluded if they had an abnormal findings on base line endoscopy (endoscopic score > 1, see below).

Study Design

The study was conducted using a randomized, active and placebo controlled, parallel-group design, with a blind evaluator.

In the previous three weeks the evaluation of the inclusion and the exclusion criteria was carried-out. It included anamnesis, physical exam, ECG, blood analysis (haematology, biochemistry and serologies for hepatitis B and C viruses, and for HIV), a 13 C-urea breath test for Helicobacter pylori (UB Test. Otsuka Pharmaceutical, S.A.), urine tests for abuse drugs, and pregnancy test in women. During the 24 hours previous to treatment an endoscopic examination to rule-out previous pathology was performed.

Subjects were randomly assigned, in a 2.5 : 2.5 : 1 ratio, into three groups that received one bottle of triflusal oral solution (600 mg) daily, two triflusal capsules (600 mg) once daily, and two placebo capsules once daily, respectively. Randomization was performed from a computer-generated balanced list based on the above-mentioned distribution. The study medications were administered daily in the Clinical Pharmacology Department of the Hospital de la Santa Creu i Sant Pau (Barcelona) during seven consecutive days. The first six days being taken at 8 a.m. and the last day (day seven) at 10 a.m. Both capsules and oral solution were taken with 220 ml of water.

Before study commencement all participants gave informed written consent, and approval was obtained from the Ethical Committee of the Hospital de la Santa Creu i Sant Pau (Barcelona) as well as from the Spanish Health Authorithies.

Assessment

Endoscopies were performed after 4-8 hours of the last dose of study drugs, using an Olympus or Pentax video gastroscope ( Olympus; Europe, Hamburg Germany). All subjects received sedation with midazolam ( dose 2 mg) or propofol. (doses ranged between 70 to 260 mg) immediately before endoscopic intubation. The esophagus, the entire stomach and duodenum were systematically examined in a proximal to distal manner. Each endoscopic procedure was documented by photographies. Hemorrhagic and erosive mucosal lesions were graded using the standard score scale devised by Lanza [START_REF] Lanza | Endoscopic evaluation of the effects of aspirin, buffered aspirin, and enteric-coated aspirin on gastric and duodenal mucusa[END_REF] and modified by the Gastroenterology Department of the Hospital de la Santa Creu i Sant Pau (Barcelona) . Scores were assigned as follows: normal, 0 points; erythema, 1 point; 1-3 erosions or petechiae, 2 points; 4-5 erosions or petechiae, 3 points; 6-10 erosions or petechiae, 4 points; > 10 erosions or petechiae, 5 points; ulcer: 6 points. Separate endoscopic injury scores were assessed (assigned) for the esophagus, stomach and duodenum until second portion. The median scores for each of the three treatment groups were calculated. Gastroscopies were documented with photographs. Two endoscopists (SS and J Ba) were involved in the study. In each volunteer basal and final endoscopies were performed by the same endoscopist who was blinded to the assigned treatment.

Subjects were provided of a diary card in order to register the following symptoms: heatburn, dysphagia, regurgitation, abdominal bloating, nausea and epigastric pain. Intensity of these symptoms was scored as follows: absence, 0; mild, 1; moderate, 2; severe, 3. These symptoms were also assessed by means of a visual analogic scale (VAS) of 100 mm in which 0 was absence of symptoms and 100 the poor possible symptom. Patients marked in the VAS from the day one to the day seven, after medication intake. Blood analysis (haemathology and biochemistry) and ECG were repeated at 24-72 hours of last dose of study drugs.

All reported adverse events and abnormal laboratory findings were recorded and tabulated by treatment group. Causality relationship was defined according to the World Health Organization criteria [START_REF] Meyboom R H B | Causal or casual?. The role of causality assessment in pharmacovigilance[END_REF].

Statistical analysis

Based on the results of Fiorucci et al [START_REF] Fiorucci | Gastrointestinal safety of NO-aspirin (NCX-4016) in healthy human volunteers: A proof of concept endoscopic study[END_REF], a sample of 50 subjects (25 in each one of the active treatment arms) would be required to achieve a 80% power to detect a difference of 1.65 points between triflusal oral solution and triflusal capsules in the modified Lanza scale.

Besides, a sample of 10 subjects in the placebo group was also included for comparison with the active drugs.

Main and secondary variables were analysed by means a Kruskall-Wallis test comparing the three study treatments. When differences were found, comparisons between groups were analysed by means a Mann-Whitney test. All analysis were performed with a two sided significance level of 0.05.

Results

One hundred and fifteen subjects participed in the screening visit. Forty four volunteers failed to meet selection criteria (28 of them because Helicobacter pylori was present) and were excluded from the study. The rest of 71 subjects had an endoscopy screening with normal results at base line. From these, 10 volunteers were considered as reserves and 61 (31 men) were included in the study. One subject dropped-out due to lack of compliance with the protocol and was not replaced. Sixty volunteers completed the study. Twenty five received triflusal capsules, 25 received triflusal solution and 10 volunteers received placebo. Demographic characteristics are shown in Table 1.

All endoscopies at baseline were considered normal in accordance with the inclusion and exclusion criteria and in any case the endoscopic score was > 1. When we used the modified Lanza scale, the median injury score at base line was 0 for the three treatment groups, both globally and separately in the stomach, esophagus and duodenum (Table 2). No differences were found between groups. After treatment, median global scores in the placebo, triflusal solution and triflusal capsules groups were, respectively: 0, 1, and 3 (p = 0.008 between three treatment groups, Kruskall-Wallis test; p = 0.003 for comparison between placebo and triflusal capsules, Mann-Whitney test; and p = 0.042 for comparison between triflusal solution and triflusal capsules, Mann-Whitney test). There were no significant differences between triflusal solution and placebo (p = 0.212, Mann-Whitney test). When considered separately, median scores for esophagus or duodenum were 0 in the three treatment groups, whereas in the stomach median scores for placebo, triflusal solution and triflusal capsules were, respectively, 0, 0 and 3 (p = 0.004 between three treatment groups, Kruskall-Wallis test; p = 0.003 for comparison between placebo and triflusal capsules, Mann-Whitney test; and p= 0.022 for comparison between triflusal solution and triflusal capsules, Mann-Whitney test). No significant differences were found between triflusal solution and placebo. Endoscopic scores after treatment are shown in Table 3. An example of gastroduodenal endoscopies for each treatment is shown in Figure 1.

Medians for symptoms in VAS and in the diary cards for placebo, triflusal solution and triflusal capsules were 0 in all treatment days. Thus, no differences were detected concerning these variables.

Adverse events were reported in two, nine and seven volunteers in the placebo, triflusal solution and triflusal capsules groups, respectively (Table 4). All adverse events were mild or moderate in intensity. None was serious and all subjects recovered.

Discussion

We conducted a proof of concept endoscopic and well powered study to compare the gastrointestinal safety of triflusal oral solution with triflusal capsules in healthy volunteers. As secondary objectives the gastrointestinal safety of triflusal oral solution and triflusal capsules were compared with placebo. Subjects enrolled in this study were well matched at baseline with a median injury score of 0 for the three groups of treatment. The median global score in the triflusal capsules group after seven days of treatment increased to 3, compared with an increase to 1 in subjects receiving triflusal solution and with an unchanged score of 0 in subjects receiving placebo. Therefore, our data provide convincing evidence that triflusal solution induce less visible changes in gastroduodenal mucosa than triflusal capsules in subjects without previous gastroduodenal disease. A possible explanation for these findings is that when triflusal is administered in a solid form (capsules) the disgregation of the drug into the stomach can not be uniform, and local concentrations can be particularly high. In addition, the intrinsic acidity of triflusal is high, with a pKa value of 3.0. However, when triflusal solution is administered an homogeneous distribution along the gastric mucose is guaranteed, as well as a faster absorption [16] and, more important, a solution containing the sodium salt of triflusal (as is the case), with a pH between 6 and 7, does not produce the local effects caused by the strong acidity of the solid form (not sodium salt). In a recently published study [START_REF] Riambau-Alonso | Estudio epidemiológico observacional para evaluar el manejo de los trastornos gastrointestinales en el paciente con tratamiento antiagregante[END_REF], performed in 609 patients undergoing chronic antiplatelet therapy who presented gastrointestinal disorders attributed to this therapy, mainly with aspirin , the change of the initial antiplatelet therapy for triflusal solution improved treatment tolerability and quality of life of patients.

Gastrointestinal toxicity of antiplatelet agents such as aspirin and clopidogrel are well known.

Even when administered at low doses, aspirin can cause serious gastrointestinal bleeding [START_REF] Patrono | Antiplatelet drugs. American College of Chest Physicians Evidence-Based Clinical Practice Guidelines (8 th edition)[END_REF] and the most frequently reported adverse events in both aspirin and clopidogrel groups in the CAPRIE [START_REF] Caprie Steering | A randomised, blinded, trial of clopiodgrel versus aspirin in patients at risk of ischaemic events (CAPRIE)[END_REF] study were gastrointestinal events, although the incidence was lower with clopidogrel. In an attempt to decrease the incidence of serious gastrointestinal side effects associated to aspirin and clopidogrel, proton pump inhibitors are frequently associated to these antiplatelet drugs and, in this sense, the American College of Cardiology/American Heart Association Guidelines in Unstable Angina/Non-ST Elevation Myocardial Infarction [START_REF] Anderson | ACC/AHA 2007 Guidelines for the Management of Patients With Unstable Angina/Non ST-Elevation Myocardial Infarction. A Report of the American College of Cardiology/American Heart Association Task Force on Practice Guidelines (Writing Committee to Revise the 2002 Guidelines for the Management of Patients With Unstable Angina/Non ST-Elevation Myocardial Infarction)[END_REF] recommends than in patients with a history of gastrointestinal bleeding, when aspirin or clopidogrel are administered alone or in combination, drugs to minimize the risk of recurrent gastrointestinal bleeding (e.g., proton pump inhibitors) should be prescribed concomitantly (Class I, Level of Evidence: B). A report of the American College of Cardiology Foundation/American College of Gastroenterology/American College of Cardiology recommends the administration of proton pump inhibitors in patients receiving antiplatelet therapy if gastrointestinal risk factors are present [START_REF] Bhatt | ACCF/ACG/AHA 2008 Expert Consensus Document on Reducing the Gastrointestinal Risks of Antiplatelet Therapy and NSAID Use. A Report of the American College of Cardiology Foundation Task Force on Clinical Expert Consensus Documents[END_REF]. Recently there is emerging evidence that concomitant administration of proton pump inhibitors diminishes the antiplatelet effect [START_REF] Gilard | Influence of omeprazol on the antiplatelet action of clopidogrel associated to aspirin[END_REF] and the clinical efficacy of clopidogrel [START_REF] Ho | Risk of adverse outcomes associated with concomitant use of clopidogrel and proton pump inhibibitors following acute coronary syndrome[END_REF]. For the above mentioned reasons it is convenient to dispose of alternative antiplatelet agents with a good gastrointestinal safety profile such as triflusal.

An antiplatelet drug in oral solution form could be useful to administer through a nasogastric tube as can be the case of patients with ischemic stroke [START_REF] Adams | Guidelines for the early management of adults with ischemic stroke[END_REF]. In this case an antiplatelet drug in form of oral solution could be of interest. On the one hand it avoids the process of trituration of a solid antiplatelet drug, which is an alteration of the original pharmaceutical form, and on the other hand, it spares time to the patient's careers. Moreover, an oral solution can be also useful to be administered to patients with dysphagia to solids for any reason but needing treatment with antiplatelet drugs. In this sense, it is known that about 50% of institutionalized elderly people has swallowing disorders [START_REF] Finiels | Les troubles de la déglutition du sujet âgé[END_REF].

Our study has some limitations that should be commented. Firstly, it should be taken into account that changes in gastroduodenal mucose in this kind of studies are not necessarily good predictors of major upper gastrointestinal complications. In this sense, it should me mentioned that results from several endoscopic studies carried out in healthy volunteers supported the hypothesis that enteric-coated aspirin caused less gastric erosion and microbleeding than regular formulations [START_REF] Hotfiezer | Comparison of the effects of regular and enteric-coated aspirin on gastroduodenal mucosa on man[END_REF][START_REF] Lanza | Endoscopic evaluation of the effects of aspirin, buffered aspirin, and enteric-coated aspirin on gastric and duodenal mucosa[END_REF][START_REF] Hawthorne | Aspirin-induced gastric mucosa damage: prevention by enteric-coating and relation to prostaglandin synthesis[END_REF] but no clinical benefits in terms of reduction of gastrointestinal bleeding or ulceration with enteric coating have, therefore, been successfully demonstrated, although the endoscopic studies show that potentially these benefits could exist [START_REF] Walker | Does enteric-coated aspirin result in a lower incidence of gastrointestinal complications compared to normal aspirin?[END_REF].

Secondly, the sample size was calculated from results of a previously published study [START_REF] Fiorucci | Gastrointestinal safety of NO-aspirin (NCX-4016) in healthy human volunteers: A proof of concept endoscopic study[END_REF] whereas the assessment of hemorrhagic and erosive mucosal lesions were grated using the standard score scale devised by Lanza [START_REF] Lanza | Endoscopic evaluation of the effects of aspirin, buffered aspirin, and enteric-coated aspirin on gastric and duodenal mucusa[END_REF] and modified by the Gastroenterology Department of the Hospital de la Santa Creu i Sant Pau (Barcelona) . It is not the most desirable from a methodological point of view, but endoscopists participating in the study preferred the scale that they used daily to assess endoscopic results. In third place, the novel galenic presentation of triflusal improved the topical gastroduodenal toxicity but if the lower endoscopically apparent mucosal damage of triflusal solution versus triflusal capsules signifies less major upper gastrointestinal complications remains to be elucidated.

Global tolerability of both triflusal solution and triflusal capsules was good, with no serious adverse events reported.

In conclusion, in subjects without previous gastroduodenal disease, triflusal solution induced less endoscopically apparent mucosal damage than triflusal capsules and did not induce more damage than placebo. Further studies will be required to confirm the long-term safety of triflusal solution in patients.

Figure 1 .

 1 Figure 1. Photography samples obtained during endoscopic in each treatment group ; A) normal mucosa in stomach fundus after treatment with placebo ; B) normal mucosa in stomach after treatment with triflusal solution; and C) mild erosive mucosal lesions, grade ? after treatment with triflusal capsules.

Scientific Research Program of the Spanish Ministry of Science and Technology.

Acknowledgements

The authors thank J. Uriach y Compañía, S.A. (Palau-solità I Plegamans, Barcelona, Spain) for financial support to this study, and to Montserrat Puntes, Joan Martínez and Salvador Rico for their assistance during data collection. This study was partially supported by the National

Conflict of interest

Iñaki Izquierdo, Javier Borja and Iris Blanch are employees of J. Uriach y Compañía, S.A.

Esther Donado was an employee of J. Uriach y Compañía, S.A. when the study was performed. Manel Barbanoj has been external advisor of J. Uriach y Compañía, S.A.

Treatment

Esophagus Stomach Duodenum Global

Placebo 0 0 0 0 (0,1) (0,1) (0,0) (0,1)

Triflusal solution 0 0 0 0 (0,0) (0,1) (0,0) (0,1)

Triflusal capsules 0 0 0 0 (0,1) (0,1) (0,0) (0,1)

No differences were found between groups.


Enzo Weber
email: enzo.weber@wiwi.uni-regensburg.de

C32 -Time-Series Models < C3 -Econometric Methods: Multiple/Simultaneous Equation Models < C -Mathematical and Quantitative Methods

Keywords: Economic Integration, Cointegration, Common Cycles, South-East Asia Economic Integration, Cointegration, Common Cycles, South-East Asia JEL classification: E32, F15, C32

The subject of this paper tackles macroeconomic integration of the South-East Asian countries South Korea, Singapore and Taiwan. Economically, the analysis is based on notions of stochastic long-run convergence and business cycle synchrony in the GDPs.

According tests for cointegration and common serial correlation features reveal a high degree of coherence in long-run growth and medium-run fluctuations. This allows extracting a common stochastic growth trend and a common business cycle. Further analysis shows that both of these components are subject to stronger influences from the US than from Japan. Convergence towards these matured economies conspicuously appears since the 1990s.

Introduction

The last decades of economic development in South-East Asia are marked by vibrant dynamics as the region has witnessed extraordinary growth, which did not go without striking inequality and severe crises. A recent example has been the Asian financial crisis in 1997/98, which occurred in a period of booming economy and deepening interaction between the region's countries. While some nations could make considerable progress, others did far less participate in the dynamic development process. As matter of fact, the members of the main regional organisations, ASEAN and APEC, range from highly industrialised to developing countries, the integration concepts from "open regionalism" to discriminatory trade policies. Thereby, the most successful industrialising economies have soon been called the "Asian Tigers".

From the scientific point of view, it remains an open task to examine the ambiguous character of the South-East Asian economies, even after some work, mainly done on shock identification and income convergence, has shed some light on this topic. For instance, [START_REF] Bayoumi | One Money or Many? Analyzing the Prospects of Monetary Unification in Various Parts of the World[END_REF] find evidence for regional clustering in terms of demand and supply innovations, which bears a certain scope for monetary cooperation. [START_REF] Zhang | Is a monetary union feasible for East Asia[END_REF] and [START_REF] Bacha | A common currency area for ASEAN? issues and feasibility[END_REF] confirm such coherence in certain subregions, while East Asia as a whole is characterised by strong heterogeneity. [START_REF] Kwan | Economic Interdependence in the Asia-Pacific Region: Towards a Yen Bloc[END_REF] creates the notion of a "flying geese pattern" for describing the way the emerging Asian economies sequentially started their catching-up processes. In this line of research, [START_REF] Lee | Income convergence between Japan and the rest of East Asian countries[END_REF] can establish real long-run convergence of Singapore towards Japan, whereas the remaining ASEAN nations failed in catching up. [START_REF] Lim | Convergence and catching up in ASEAN: a comparative analysis[END_REF] find mixed evidence for convergence within the ASEAN group.

Surely, the special regional scenery with its varying economic conditions and political historical backgrounds raises several questions: Most profoundly, does South-East Asia justify using the notion of one economic region? In detail, do the different countries share common macroeconomic characteristics, and which is the nature of transmission effects between them? Seen from a broader perspective, has the regional development led to convergence towards the industrialised world? And at last, which is the importance of foreign economic influences on South-East Asia? Answering these questions would not only give a more profound insight into the functioning and structure of the world's most dynamic and forward-looking region. On top, it would provide important information for understanding the difficulties in enhancing sustained progress and ensuring macro-stability despite the high vulnerability, which has This paper approaches the outlined problems in the context of an empirical time series analysis, proceeding on an aggregated macro-level. The examination makes use of recent econometric advances in the field of common features (e.g. Engle and Kozicki 1993, [START_REF] Hecq | Common Cyclical Features Analysis in VAR Models with Cointegration[END_REF]). Namely, the analysis of common stochastic trends and common cyclical dynamics provides adequate means for the given target of finding evidence about stance and course of a macroeconomic convergence: First, relying on the [START_REF] Bernard | Convergence in international output[END_REF] concept of income convergence, cointegration is a natural prerequisite for similar growth paths in the long run. Second, synchronous business cycles indicate common grounds mainly in the medium-frequency macroeconomic processes. A particular contribution of the underlying paper consists in analysing both the growth and cycle side of the economy in a unified methodological framework.

The first step of the analysis consists of testing for cointegration and common cycles in the GDPs of South-East Asian countries. Thereby, the exceptionally successful Tigers South Korea, Singapore and Taiwan exemplify the impressive emergence of economic progress and might represent one cluster as identified by [START_REF] Bayoumi | One Money or Many? Analyzing the Prospects of Monetary Unification in Various Parts of the World[END_REF], [START_REF] Zhang | Is a monetary union feasible for East Asia[END_REF] or [START_REF] Bacha | A common currency area for ASEAN? issues and feasibility[END_REF]. Based on the obtained results, a common growth trend and a common business cycle are filtered out of the data. These series allow interesting insights into long-and medium-run connections to the industrialised world: While considerable impacts from the US become manifest, evidence for convergence strikingly rises in the mid-1990s. The last point gives an answer to the problems in the existing literature of finding support for the time series convergence hypothesis. The proceeding differs from the main existing research directions, which focus for example on the identification and comparison of national supply and demand shocks [START_REF] Bayoumi | One Money or Many? Analyzing the Prospects of Monetary Unification in Various Parts of the World[END_REF], or cross-section panel convergence; for the latter, see for example [START_REF] Engelbrecht | Economic Growth and Convergence Amongst the APEC Economies 1965-1990[END_REF], who characterise the APEC countries as a convergence club.

Preceding the empirical part, section 2 supplies the foundation of the economic concept consisting of theories on convergence, business cycle transmission, growth and optimal currency areas. In the section thereafter, the econometric methods, cointegration, common cycle features and permanent-transitory decomposition, are to be introduced. Section 4 presents the empirical results for the South-East Asian region and its outward relations. Finally, the last section evaluates and summarises the outcome of the econometric investigation. Before discussing and applying the econometric methods in the next sections, it is worthwhile to briefly consider the economic notion that underlies the following analysis. As has been mentioned in the introduction, the examination aims at grasping the actual stance of regional economic integration in South-East Asia. My concept behind the term integration is built on stochastic trending behaviour of the national GDPs in the long run as well as cyclical fluctuations in the short and medium run. The word cycle has not necessarily to be taken literally, but rather in a sense of describing transitory movements in contrast to permanent trending (e.g. [START_REF] Vahid | Common Trends and Common Cycles[END_REF]. Of course, this definition still comprises the original meaning of regular periodical up-and downswings in a macro economy.

The motivation for this concept may be best illustrated by the theory of optimal currency areas. As a core result, it recommends the introduction of a common currency representing a very high stage of integration, if different exchange rates and separated monetary policies have lost their necessity. Above all, this is the case when synchronous cycles facilitate adequate reactions by one common central bank, the foreign exchange is not confronted with a variety of idiosyncratic national shocks and a long-run growth equilibrium between the economies provides preconditions for development in a unified economic system.

Focusing on the latter point, the neoclassical growth theory implies that the steadystate of per capita output does not depend on initial conditions. In detail, diminishing marginal returns are the base for higher growth prospects in countries with lower current income levels, at least theoretically. As a consequence, a common convergence target between different economies is determined in the long run. Only in case of structural microeconomic parameters differing across countries, stable gaps between outputs would persist due to distinct production functions.

In a stochastic framework, [START_REF] Bernard | Interpreting tests of the convergence hypotheses[END_REF] require for convergence that the gap in outputs between country 1 and 2 is closed while increasing the forecast horizon:

lim k→∞ E t (y 1,t+k -y 2,t+k ) = 0 . (1)
Since this definition rules out persistent shocks to the output gap, it implies that per capita outputs are in pairs cointegrated with the vector (

Model Framework

The basic data generating process in the econometric procedure shall be approximated by a VAR with lag length q + 1

y t = c 0 + c * 1 t + c * 2 b t + c 3 d t + q+1 i=1 A * i y t-i + u t , (2)
where y t contains the n endogenous variables (GDPs), A * i are n×n coefficient matrices and u t is an n-dimensional vector of white noise errors. The deterministic terms are constant, linear trend (t), level breaks (b t ), which might accomodate economic disruptions, and impulse dummies (d t ), which guarantee normality of the residuals.

Given the presence of unit roots in the per capita GDPs, according to [START_REF] Johansen | Likelihood-based Inference in Cointegrated Vector Autoregressive Models[END_REF], the commonness of nr stochastic trends is reflected by a reduced rank of A * (1), with

A * (L) = I n -q+1 i=1 A * i L i .
Consequently, one can write A * (1) = -αβ , where β spans the space of the r cointegrating vectors, and α contains the corresponding adjustment coefficients. Granger's representation theorem leads to the VECM

∆y t = α(β y t-1 + c 1 (t-1) + c 2 b t-1 ) + c 0 + c 3 d t + q i=1 A i ∆y t-i + u t , (3)
with A i = -q+1 j=i+1 A * j , i = 1, . . . , q. This representation assumes that trend and breaks are absorbed in the cointegrating relation. Note that in (3) lagged intervention dummies, which condition the likelihood function in each subsample (defined by the break dates), as in [START_REF] Johansen | Cointegration analysis in the presence of structural breaks in the deterministic trend[END_REF], are not displayed for simplicity. For illustration, in the bivariate case the real income convergence would imply β = (1, -1) in (3).

Trend Analysis

As the basis for long-run growth analysis, the unit root behaviour of (non-breaking) series is checked by ADF tests, with constant, trend and centred seasonal dummies included.

The lag length is set following the usual information criteria and autocorrelation tests. Various authors found that the presence of structural breaks distorts the unit root test results, see i.e. [START_REF] Perron | The great crash, the oil price shock and the unit root hypothesis[END_REF]. Certainly, there is no doubt that such shifts have occurred in the Asian financial crisis. Here, I follow [START_REF] Saikkonen | Testing for a unit root in a time series with a level shift at unknown time[END_REF], who propose first estimating the deterministic nuisance parameters and afterwards testing the residuals for non-stationarity. Accordingly, in the first step a GLS regression of the time series on constant, trend, dummies and shift is run. As in the case of the Asian crisis the shift dates are evident, I assume the break points to be known; nevertheless, endogenous break point search does not alter the basic conclusions. In the second step, an ADF-type test on the estimated residuals is performed. For critical values of the t-statistic and additional correction terms in the regression see [START_REF] Lanne | Comparison of unit root tests for time series with level shifts[END_REF]. [START_REF] Johansen | The role of the constant and linear terms in cointegration analysis of nonstationary time series[END_REF][START_REF] Johansen | Likelihood-based Inference in Cointegrated Vector Autoregressive Models[END_REF] provides a test for cointegration in the VECM from (3), Johansen et al. ( 2000) incorporate structural breaks. Their likelihood ratio (LR) trace test statistic for the null hypothesis of at most r cointegrating relations is given by

Λ(r) = -T n i=r+1 log(1 -λi ) , (4)
where n is the number of endogenous variables and T the number of observations. λi denotes the i-th largest squared sample canonical correlation between ∆y t and the respective cointegrating relation, both corrected for the influence of the remaining regressors. Since the trace test is known to be distorted in small samples, I implement a correction of the test statistic based on [START_REF] Cheung | Finite sample size of Johansen's likelihood ratio tests for cointegration[END_REF]. Critical values are obtained by computing the response surfaces in Doornik (1998), or [START_REF] Trenkler | Determining p-values for systems cointegration tests with a prior adjustment for deterministic terms[END_REF] in case of breaks. Given the cointegration results, further LR tests are applied to check the null hypotheses of various parameter restrictions against the unrestricted model.

Cycle Analysis

Besides cointegration as rank reduction in the long-run multiplier matrix, I concentrate on short-and medium-term business cycle synchrony implying reduced rank in the short-run dynamics. More precisely, this means looking for a linear combination of the differenced time series, which is not predictable from the relevant past, thus eliminates all systematic autocorrelation structure.

Engle and Kozicki (1993) introduced the common cycles feature, which was taken up by [START_REF] Vahid | Common Trends and Common Cycles[END_REF]. They define a cycle as common, if some linear combination of the autocorrelated series is not serially correlated itself. This implies that there exists an n × s matrix γ pre-multiplied in equation ( 3), such that γ α = 0 and γ A i = 0 for i = 1, . . . , q. That is, if γ lies in the intersection of the left nullspaces of the longand short-run adjustment coefficients, one gets the innovation process γ ∆y t = γ u t .

Integration of both sides clarifies that γ removes as well the cyclical component in the levels y t , leaving a random walk with naturally uncorrelated innovations. It is obviously not worth considering VECMs with q = 0, because γ would simply be the orthogonal complement of α.

Testing the null hypothesis of at most s common cycle vectors is done by canonical correlation analysis between ∆y t and the regressors from (3), ∆y t-i (i = 1, . . . , q) and β y t-1 . Note that s cannot exceed nr (Vahid and Engle 1993), because the cofeature matrix must lie in the space of the n × (nr) orthogonal complement of α. The canonical correlations are conditioned on a constant and outlier neutralisation dummies. With the eigenvalues η i (in ascending order), the LR test statistic for the null hypothesis of at least

s common cycle vectors CC(s) = -(T -n(q -1) -r) s i=1 log(1 -ηi ) (5)
is calculated. The relevant χ 2 -distribution has s(r + nq)s(ns) degrees of freedom.

This can be seen, when applying the common cycle restrictions to the VECM from (3):

I s γ 0 (n-s)×s I n-s ∆y t = 0 s×r α (β y t-1 +c 1 (t-1)+c 2 b t-1 )+c * 0 +c * 3 d t + q i=1 0 s×n Ãi ∆y t-i +v t . (6)
The cofeature matrix has been normalised such that γ = (I s : γ ). α and the Ãi contain the lower ns rows of α and the A i ; the upper rows are filled by s•r zero restrictions on α and s•n on each of the q A i . c * 0 and c * 3 are given by the cofeature matrix multiplied by c 0 and c 3 . The efficient estimation of the cofeature matrix requires s•(n-s) additional coefficients, leaving a parameter difference equal to the above-mentioned number of degrees of freedom.

From [START_REF] Cheung | Finite sample size of Johansen's likelihood ratio tests for cointegration[END_REF] it is seen that the transitory movements of s differences are absorbed in the cofeature relations. This implies that the s smallest canonical correlations should be tested insignificant by [START_REF] Canova | Trade interdependence and the international business cycle[END_REF].

While the common cycles concept calls for linear combinations not predictable at all, one could reduce this requirement to the short-run, meaning high frequencies. This leads to 7 allowing for different factors generating the long-and short-term dynamics as in [START_REF] Hecq | Common Cyclical Features Analysis in VAR Models with Cointegration[END_REF]. In detail, the low-frequency fluctuations from the adjustment processes are considered as idiosyncratic. Therefore, in the search for unpredictable linear combinations, one corrects for the cointegration equilibrium effects: γ (∆y tαβ y t-1 ) = γ u t . Besides increasing flexibility, the high-frequency variant has the advantage of limiting the number of common cycle vectors only to n -1 instead of nr. The appropriate model differs from [START_REF] Cheung | Finite sample size of Johansen's likelihood ratio tests for cointegration[END_REF] only in that α is left unrestricted. Accordingly, the test statistic is the same as in [START_REF] Canova | Trade interdependence and the international business cycle[END_REF] with r set to zero. The eigenvalues are estimated in the same canonical correlation programme, though now additionally conditioned on the error correction terms.

Permanent-Transitory Decomposition

Until this point, I have presented tests for the number of common trends and common cycles among several time series. These numbers determined, the trending and cyclical components can be filtered out of the data. In this, I follow [START_REF] Proietti | Short-Run Dynamics in Cointegrated Systems[END_REF], who provides the permanent-transitory decomposition, and [START_REF] Hecq | Permanent-Transitory Decomposition in VAR Models with Cointegration and Common Cycles[END_REF], who add the common cycles feature.

First, define the matrix

P = (A(1) -αβ ) -1 α(β (A(1) -αβ ) -1 α) -1 β , (7)
where A(L) = I n -q i=1 A i L i . Then, the common trends are given by

(I -P )(A(1) -αβ ) -1 α ⊥ (α ⊥ α ⊥ ) -1 α ⊥ A(L)y t , (8)
with ⊥ denoting the orthogonal complement (thus α α ⊥ = 0, where both α and α ⊥ have full column rank). The common cycles result as

γ⊥ (γ ⊥ γ⊥ ) -1 γ ⊥ (P y t -(I -P )(A(1) -αβ ) -1 Ã(L)∆y t ) , (9)
where Ã(L) = q-1 i=0 Ãi L i and Ãi = q j=i+1 A j , i = 0, . . . , q -1. Note that in case of highfrequency common cycles, expression [START_REF] Engelbrecht | Economic Growth and Convergence Amongst the APEC Economies 1965-1990[END_REF] does not contain the transitory components corresponding to the equilibrium adjustment processes, which remain idiosyncratic. Because of inflation, different currencies and price levels as well as different population sizes, the raw GDP series are no adequate data. Accordingly, the series have been transformed as follows: Per capita GDP has been calculated dividing by total population, which was linearly interpolated to gain quarterly values. The nominal data have been deflated to the year 2000 level using the GDP price deflator. Finally, the year 2000 purchasing power parity conversion factors from the international comparison program of the World Bank have been employed to transform all series into US dollar. 1 The calculated variables can be interpreted as per quarter amount of dollars one would have needed in the USA in 2000, to reach the same living standard as the foreigner in the respective country and period.

The Census X-12 seasonally adjusted GDPs are displayed in Figure 1. A first graphical inspection brings the following stylised facts to the fore: The South-East Asian countries seem to share a regionally balanced growth path, multiplying their GDPs within the last decades. The temporary halting of this successful development process due to the 1997/98 financial crisis is clearly visible in the series of Korea and Singapore, countries known for having struggled the most at the time. The subsequent upswing was interrupted by the world economic recession, which can as well be seen in the US series. In Japan, the economic boom in the late 1980s is traceable just as the long period of deflationary 1 For Taiwan, the factor has been calculated by a PPP update based on the 1990 relative price from Penn World Table. The trace test statistic ( 4) is employed to uncover the number of cointegrating relations r among the three series. H 0 : r = 0 is rejected on the 1% level (Λ(0) = 53.73, critical value = 48.87), and since Λ(1) = 26.17 still exceeds the 5% critical value of 25.73, there is even evidence against H 0 : r = 1. Therefore, I allow for two cointegrating relations, both of which are found clearly stationary when applying ADF tests. This result does not depend on the particular deterministic specification and sample choice, but is not robust against substantially reducing the lag length. However, as mentioned above, a small q does not appropriately capture the serial dynamics, supporting the sensibility of the current model set-up. This is underlined by Table 2, showing that the LM hypotheses of no autocorrelation up to the respective lag and the Jarque-Bera hypothesis of normality cannot be rejected.

LM

Cointegration and Convergence

The trace test results meet the Bernard and Durlauf requirement of pair wise cointegration, implying one common stochastic trend in the GDPs of all three countries. Therefore, seen from a growth perspective, the driving forces of economic development processes are shared in the South-East Asian region. As argued in section 2, this strong result can arise from common factors as well as interaction within the region. Both potential sources will be assessed as the analysis proceeds. However, the convergence definition is not limited Consequently, I test the hypothesis that the row vectors in β are equal to (1, 0, -1) and (0, 1, -1), where the variables KOR, SGP and TWN are simply in alphabetical order.

A LR test (Johansen 1995) produces a p-value of 0.37 (test statistic = 1.98 with two degrees of freedom), so that the restrictions cannot be rejected. The error correction part of model ( 3) takes the form

    ∆KOR t ∆SGP t ∆T WN t     =       -0.
(t-1)    +. . . , (10)
with standard errors in parentheses and significant estimates in bold. The deterministics reveal a significant initial gap between the Singaporean and Taiwanese GDPs, as can already be seen from Figure 1. 2 The significant linear trend in favour of the Korean compared to the Taiwanese development is mostly owed to the sustained influence of the 2001 recession and the delayed crisis impacts in Taiwan. In general, since magnitudes are rather small and simple extrapolations into the future are surely not appropriate, one should not put too much weight on the deterministic model part.

Concerning the adjustment processes, the finding is that of interdependence in the long run: Deviations from the growth equilibrium affect the further GDP development of all countries. These adjust in the "correct" direction, thereby re-equilibrating the system as time passes by. The smallest coefficients show up in the equation of Taiwan, which thus maintains the highest degree of independence. In contrast, the small economy of Singapore reveals the strongest reactions; significantly, this size effect will reappear below.

Business Cycle Coherence

Until now, the stochastic trend analysis has been subject of interest. Laying the focus on medium-run fluctuations, this section presents results on common cycles as the second conceptual element. The tests for common cycles are conducted as described in section The restriction of one high-frequency common cycle can now be applied to the VECM from

(3), in order to estimate the cofeature coefficients with standard errors. The cofeature matrix results as

γ =    1 0 -0.540 (0.125) 0 1 -1.571 (0.218)    . ( 11
)
For instance, the estimates imply that the transitory fluctuations of Korea (the first variable) on the right hand side of ( 6) can be eliminated by subtracting 0.540 times the Taiwanese variable (third). An estimate of -1 would indicate coherent cycles of equal per capita strength. Therefore, the common cycle is strongest in the Singaporean GDP, followed by Taiwan and Korea. The clear significance of the parameters supports the decision of introducing common feature restrictions.

The presence of a common business cycle imposes the question of transmission effects between the countries involved. For the high-frequency cycle, this means looking for spillover interaction that appears apart from the long-run growth effects. A logical procedure thus consists of testing Granger non-causality restrictions on the short-run dynamics in the VECM from [START_REF] Bernard | Convergence in international output[END_REF]. For this purpose, I define the hypotheses that each respective country is not subject to any short-run influences from the remaining two, or formally, that both off-diagonal elements in the respective row of all A i (i = 1, . . . , 6) equal zero. Table 3 3 Remember that s now denotes the number of high-frequency common cycle vectors.

13 Evidence against the hypothesis of no influence received is highest for Singapore, followed by Taiwan and Korea. Again, an appealing interpretation can therefore link the degree of independence from cyclical spillovers to the size of the economies. Note as well that the same order of countries was found concerning the strength of the business cycle. Naturally, the outcome of Granger causality tests depends on the set of variables included; while the system restricted to South-East Asian economies already provided interesting insights, the roles of Japan and the US as world factors will be explored in section 4.6.

F

Decomposing GDPs

The preceding sections have dealt with examining the strength of economic integration.

Thereby, high coherence could be established within South-East Asia concerning long-run growth paths as well as business cycle fluctuations. In the following, I filter out both common trend and common cycle, in order to analyse outward economic linkages of the region as a whole.

The common trend is calculated as in [START_REF] Doornik | Approximations to the asymptotic distributions of cointegration tests[END_REF], where all loadings are equal due to the restrictions on the cointegrating vector. The deterministic parts are handled as follows: First, the derived common trend is subtracted from each GDP series. The three rest components are then regressed on a constant and a linear trend. The mean constant and mean linear trend from these regressions are finally added to the common stochastic trend. This procedure takes the deterministics in the estimated VECM into account and provides a sensible measure of the regional growth path. The common cycle series is constructed from [START_REF] Engelbrecht | Economic Growth and Convergence Amongst the APEC Economies 1965-1990[END_REF]. In equivalence to the stochastic trend, constant and linear trend from an OLS regression are subtracted from the cycle. Figure 2 plots the obtained series.

Due to the calculation of the deterministics, the common stochastic trend lies midst of the underlying GDP series. In the cyclical pattern, one can detect an upswing in the early 1980s ended by the mid-80s economic crisis, a trough corresponding to the world recession in the early 1990s, the enormous recovery process after the Asian financial crisis The GDPs of Japan and the US are decomposed in comparable univariate procedures:

At first, the ARIMA models

∆JP N t =
∆U SA t-2 + v2t (13)
are estimated. Thereby, the specifications based on information criteria and parameter significance are checked by testing for absence of residual autocorrelation. Then, the obtained residuals v are sequentially accumulated to random walk series. After the deterministics have been subtracted and added as above, at last the cycle is defined as the difference between series and trend. Other sensible ARIMA settings leading to results substantially different from the following could not be found.

South-East Asia in the World Economy

Since all permanent and transitory components are derived, this section can address international linkages of the South-East Asian region. Starting with business cycle connections, for reasons of clarity the analysis is focused on the major cases of Japan and the US. In a first step, I consider the degree of alignment between the transitory fluctuations. Correlation of cycles might be due to influences in one or both directions, but can as well appear in presence of common shocks. Granger causality analysis is likely to provide further information on this topic. For this reason, in analogy to Figure 3 The left panel clarifies that systematic influences between the South-East Asian and the Japanese business cycles do not exist. This result is not due to low test power in the short time windows, since longer estimation periods neither allow to reject the null hypotheses. In contrast, evidence for US influences on the South-East Asian cycle can be found in the right panel. These are however restricted to the periods of high correlations, which had been identified above. Nonetheless, it can be stated that cyclical coherence is a result of single-direction US impacts, while Japanese spillovers are not apparent.

Obviously, despite the regional proximity, Japan cannot match with the super power US in originating major business cycle impulses.

The second point of interest is the stance of the South-East Asian region in international growth and convergence processes. Again, this issue is addressed following the Bernard and Durlauf concept, which suggests testing for cointegration between the common South-East Asian trend and the permanent components of Japan and the US. Since rolling windows normally provide too few observations for this type of analysis, the trace test statistics in Figure 5 are estimated backward recursively, meaning samples with fixed end and moving starting point. By construction of the stochastic trend series, the lag length q is zero, and constant and linear trend have been included as usual. For Japan, the null hypothesis of no cointegration can be rejected from 1993:2 onwards.

The US statistic reveals far less significance, but reaches the critical value in 1998:1.

Even as this date seems to ascribe a decisive effect to the Asian financial crisis, neither including a structural break nor cutting the sample in 1997:4 can reveal cointegration in earlier periods. That is, data is consistent with long-run linkages between South-East Asia and the two matured industrial nations emerging in the last one or two decades.

Problems of time series tests in finding convergence can thus be attributed to dynamic adjustments of the structural growth properties through time; this finding corresponds to the protectionist and regulatory course of the Tigers after World War II, which gave way to market-and trade-oriented strategies later on.

In the VECMs, which are estimated for the time spans indicated by the earliest significant trace tests, the hypothesis β = (1, -1) can be checked. For Japan, the LR p-value is 0. In the deterministics, the significant constants mirror the considerable initial lead of Japan and even more the US over South-East Asia. However, this gap is gradually reduced, as can be seen from the significant linear trends. Thereby, the speed of catching-up towards Japan (21.66 per capita PPP US $ per quarter) nearly doubles the one towards the US (12.63), a consequence of the Japanese "lost decade". In sum, all the convergence prerequisites, cointegration, equal weights and catching-up, are fulfilled.

From the adjustment vectors it can be seen that long-run inter dependence prevails between South-East Asia and Japan, even if the higher influence naturally comes from the latter. Contrastingly and in line with a priori expectations, the US represents a weakly exogenous variable, inducing substantial adjustment of South-East Asia. Including all three GDP trends in a common system confirms this constellation: The far largest proportion of long-run variation in South-East Asian growth turns out to be governed by US shocks.

Finally, it might be worthwhile shortly considering the special role of the fourth tiger Hong Kong: Cointegration tests and VECM estimations remained unclear through several specifications, detailed results are left out to save space. When included in the South-East

Asian system, adjustment of Hong Kong to the cointegrating relations either did not reach significance, or proved even wrong-directed, thereby enlarging occurring equilibrium deviations. Since stability of the system is nonetheless guaranteed by the neighbour countries, this constellation could possibly be interpreted as favouring a regional leading role of Hong Kong. While this may be explained by extreme openness in goods and financial markets, tight contacts to the US and China, and extremely fast recovery processes in the aftermath of economic crises, further research is required to substantiate a consistent argumentation.

Concluding Summary

In this paper, I have focused on stance and properties of the macroeconomic integration in the South-East Asian region. Stochastic convergence, business cycle coherence as well as intraregional and foreign impulse transmission were the main points of interest. In summarising the empirical results, I will conclude by closing the loop to the questions set out in the introduction.

The cointegration analysis has been motivated by a convergence criterion arising from neoclassical growth theory, which requires a cointegrating long-run equilibrium with equal weights on different per capita GDPs. In fact, evidence is in favour of common stochastic trends in the GDPs of South Korea, Singapore and Taiwan, which convincingly meet the convergence conditions. As the second element in the integration concept, I examined the synchrony of business cycle fluctuations. In this, a cycle is defined as common, if it is possible to cancel out all autocorrelation dynamics in a linear combination of the GDPs.

Applying a high-frequency common cycles test, the null hypothesis of a single common serial correlation feature could not be rejected. Spillovers of growth shocks emerge in the domains of both the long and the short run and are consistent with the different size of the economies.

It turns out that in terms of fundamental macro developments, the three countries form a homogenous economic region. Important implications relate to various topics: First, openness in goods and financial markets has brought all countries on an obviously successful growth path. Future trade arrangements and FDI policies can surely take advantage of this fact. Furthermore, the high degree of homogeneity is likely to facilitate regional monetary cooperation, because similar business cycles and structural growth determinants prevent the problem of inappropriate common policies in presence of diverging national developments. Thereby, the task of exchange rate arrangements is of particular importance in view of Asian structural vulnerability, which might be reduced in a more robust and credible currency bloc. In general, regional institutionalisation is straightforwardly tackled both as outcome of and as prerequisite for a sound integrated development.

Based on the test results, the common stochastic trend and cycle were extracted from the data, representing the South-East Asian region in a broader analysis: In detail, the focus is on examining the relations towards the leading economies of Japan and the US.

Periodical correlations of the business cycles were found only with the US, which prove causal for the South-East Asian fluctuations. Long-run convergence towards Japan can be confirmed since the early 1990s and towards the US since the late 1990s. While the latter again comes out as an exogenous driving force of growth processes, mutuality characterises the connections to Japan.

These pieces of evidence raise two central issues: On the one hand, the newly industrialised South-East Asian countries profoundly head towards full integration into the group of developed economies, thereby setting an example of successful growth strategies. On

  threat in the past. In this context, enhanced cooperation based on deepening integration could bear high potential for stabilisation policies.

  Simulated critical values for the null hypothesis of non-stationarity are taken from Davidson and MacKinnon (1993).

  This study employs GDPs of the Asian Tigers South Korea ("Korea" in the following), Singapore and Taiwan as standard measures of aggregated economic development. The fourth tiger, Hong Kong, is not included in the final model versions for reasons explained towards the end of the paper. Further South-East Asian countries like Malaysia or Thailand would be worth considering, however limited availability of quarterly GDP series of sufficient length and quality would introduce the usual severe caveats of multivariate time series analysis; nevertheless, an examination covering a larger set of Asian Pacific countries may be found in[START_REF] Weber | Macroeconomic Integration in Asia Pacific: Common Stochastic Trends and Business Cycle Coherence[END_REF]. Here, the scope is broadened by the USA as the world economic leader and Japan as the forerunner in the Asian industrialisation representing foreign influences. Quarterly data are taken out of the EcoWin databases (for Japan from IFS) for the sample 1975:1-2006:4.

9 PageFigure 1 :

 91 Figure 1: Real GDPs (2000 per capita purchasing power US $)

  , I discuss the model specification: In the deterministic part, constant and trend are included. However, any level breaks b, above all at the time around the financial crisis, were judged clearly insignificant by LR tests and proved irrelevant for the long-run model properties. Even though such permanent shifts were dispensable, three impulse dummies d neutralise extreme residual outliers: 1988:1 might be associated to a short-time growth peak following the slow-down after the "Black Monday", in 1998:1 the financial crisis effect mostly in the Korean series is accommodated, and 2003:3 corresponds to an exceptional quarter in the Singaporean and Taiwanese recovery process after the world economic downturn. The Hannan-Quinn and Schwarz Criteria preferred a short lag length of only one quarter. Since such a specification left considerable autocorrelation in the residuals, I follow the Akaike Criterion, which suggests setting q = 6.

  but comprises equal weights in the long-run relations; otherwise, the strength of the stochastic trend would differ across nations, reflecting non-aligned growth paths.

Figure 2 :

 2 Figure 2: Common South-East Asian trend and high-frequency cycle

  For this purpose, Figure3(left panel) presents rolling five-year correlations between the South-East Asian and the respective foreign cycle. The window size seems appropriate for business cycle horizons, since shorter time spans merely produce volatile and hardly interpretable correlations, while larger ones are not likely to catch the development of coherence over time. The time axis refers to the first quarter in the respective window.

Figure 3 :

 3 Figure 3: Correlations between South-East Asian and foreign cycles (see right panel)

Figure 4 :

 4 Figure 4: LR causality tests between South-East Asian and foreign cycles

  South-East Asia and JapanTrace test for South-East Asia and USA 10% critical value

Figure 5 :

 5 Figure 5: Backward recursive trace tests between South-East Asian and foreign trends

20 Page

 20 the other hand, substantial dependence on influences from abroad should be taken into account. Such a situation bears the potential of permanently initialising further progress, but latent threats in a globalising world are to be treated with caution. Future research might explore the Chinese role in the interrelations of the new Asian growth dynamics, the well-known substantial data problems.

Table 1 :

 1 Unit root tests on GDPs

		KOR	SGP TWN JPN	USA
	t-value	-0.87	-2.50	-2.16 -1.73	-2.41
	break date 1998:1 1998:1	-	-	-
	lag length	3	6	6	3	2
	10% critical values: with break -2.76, else (ADF) -3.13
		constant and trend included
	4.2 Model Specification

Convergence among the South-East Asian countries of Korea, Singapore and Taiwan is analysed within the trivariate VECM from (3). Before addressing econometric results

Table 2 :

 2 Specification tests: p-values

	1) LM(2) LM(4) LM(8) JB
	0.56	0.77	0.60	0.10	0.47

  Since two cointegrating relations have been detected, nr = 1 is an upper bound for the number of common cycle vectors. For the according hypothesis H 0 : s ≥ 1 the p-value equals 0.15 (test statistic = 24.10 with 18 degrees of freedom), so that the presence of one common cycle vector can be accepted.By the high-frequency variant, it is possible to test for the maximum number of two common cycle vectors. Since a positive result would indicate the presence of a single (high-frequency) cyclical component shared by all countries involved, this task is of special interest. With a p-value of 0.48, the hypothesis H 0 : s ≥ 1 (test statistic = 15.56 with 16 degrees of freedom) cannot be rejected.3 Even for H 0 : s ≥ 2, the p-value still comes to 0.18 (test statistic = 41.45 with 34 degrees of freedom), allowing the adoption of two common cycle vectors. Since both test variants justify adopting the minimal statistically attainable number of common transitory components, they provide evidence for close connections in the cyclical fluctuations of the different GDPs.

	3.3.
	F o r
	P
	e
	e r
	R
	e
	v i e
	w

2

Including the constants into the error correction term easily passes a LR test.

  0 : SGP, T WN → KOR H 0 : KOR, T WN → SGP H 0 : KOR, SGP → T WN

	displays the LR test statistics with 12 degrees of freedom.
	20.47	51.76 * *	27.62 *
	** , * : H 0 can be rejected at 5% respectively 1% significance level
	(5% and 1% critical values: 21.03, 26.22
	o r
	P
	e
	e r
		R
		e
		v i e
		w

H

Table 3 :

 3 LR tests on short-run Granger non-causality

  , I estimate fiveyear rolling VAR models including the South-East Asian and one of the foreign cycles, respectively. A lag length of two proved a sensible choice, which is not decisive for the qualitative results. LR tests of zero restrictions on the respective off-diagonal parameters are performed. This corresponds to the hypotheses in the legends of Figure4, which presents the rolling LR test statistics from both models.

	14		24
	12		20
	10		16
	8
			12
	6
	4 0 2	F o r 78 80 82 84 86 88 90 92 94 96 98 00	8 4 0	78 80 82 84 86 88 90 92 94 96 98 00
		H0: no Japanese influence H0: no South-East Asian influence P 5% critical value		H0: no US influence H0: no South-East Asian influence 5% critical value
		e
		e r
			R
			e
				v i e
				w

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

Acknowledgements

This research was supported by the Deutsche Forschungsgemeinschaft through the CRC 649 "Economic Risk". I am grateful to an anonymous referee, Jürgen Wolters and Cordelia Thielitz for their comments. Of course, all remaining errors are my own.