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TGFBR1*6A and Cancer Risk: A Meta-Analysis of Seven Case-Control Studies

From the Cancer Genetics Program, Division of Hematology/Oncology; the Center for Medical Genetics, Evanston Northwestern Healthcare; and the Department of Preventive Medicine, Feinberg School of Medicine, Northwestern University and Robert H. Lurie Comprehensive Cancer Center of Northwestern University, Chicago, IL; and Clinical Genetics Service and the Department of Medicine, Memorial Sloan-Kettering Cancer Center, New York, NY.

Address reprint requests to Boris Pasche, MD, PhD, FACP, Cancer Genetics Program, Division of Hematology/Oncology, Department of Medicine, and Robert H. Lurie Comprehensive Cancer Center, Northwestern University Feinberg School of Medicine, 676 N. St Clair St, Suite 880, Chicago, IL 60611; e-mail: b-pasche{at}northwestern.edu.

	ABSTRACT

TOP ABSTRACT INTRODUCTION PATIENTS AND METHODS RESULTS DISCUSSION REFERENCES

 
Purpose: TGFBR1*6A is a hypomorphic polymorphic allele of the type I transforming growth factor beta receptor (TGFBR1). TGFBR1*6A is a candidate tumor susceptibility allele that has been associated with an increased incidence of various types of cancer. This study was undertaken to analyze all published case-control studies on TGFBR1*6A and cancer and determine whether TGFBR1*6A is associated with cancer.

Patients and Methods: All published case-control studies assessing the germline frequency of TGFBR1*6A were included. Studies assessing TGFBR1*6A in tumors were excluded. The results of seven studies comprising 2,438 cases and 1,846 controls were pooled and analyzed.

Results: Overall, TGFBR1*6A carriers have a 26% increased risk of cancer (odds ratio [OR], 1.26; 95% confidence interval [CI], 1.07 to 1.49). Cancer risk for TGFBR1*6A homozygotes (OR, 2.53; 95% CI, 1.39 to 4.61) is twice that of TGFBR1*6A heterozygotes (OR, 1.26; 95% CI, 1.04 to 1.51). Analysis of various types of tumors shows that TGFBR1*6A carriers are at increased risk of developing breast cancer (OR, 1.48; 95% CI, 1.11 to 1.96), hematological malignancies (OR, 1.70; 95% CI, 1.13 to 2.54), and ovarian cancer (OR, 1.53; 95% CI, 1.07 to 2.17). Carriers of TGFBR1*6A who are from the United States are at increased risk of colorectal cancer (OR, 1.38; 95% CI, 1.02 to 1.86). However, Southern European TGFBR1*6A carriers have no increased colorectal cancer risk. There is no association between TGFBR1*6A and bladder cancer.

Conclusion: TGFBR1*6A is emerging as a highfrequency, low-penetrance tumor susceptibility allele that predisposes to the development of breast, ovarian, and colorectal cancer, as well as hematologic malignancies.

	INTRODUCTION

TOP ABSTRACT INTRODUCTION PATIENTS AND METHODS RESULTS DISCUSSION REFERENCES

 
RECENT ANALYSES of cohorts of twins show a relatively large effect of heritability for several forms of cancer, suggesting that our current knowledge of cancer genetics is limited.¹ This effect is likely due to a combination of low-penetrance tumor susceptibility genes. Such variants are relatively common in the population and may confer a much higher attributable risk in the general population than rare mutations in high-penetrance cancer susceptibility genes, such as BRCA1, BRCA2, and APC.

Candidate low-penetrance genes are usually chosen on the basis of biologic plausibility. Alterations in their protein sequence, and therefore its function, could affect pathways involved in cell-growth control, detoxification, and carcinogenesis. Several polymorphisms of genes involved in these pathways have been identified during the past few years. Transforming growth factor beta (TGF-ß) fulfills the requirements for designation as a cancer-related pathway. TGF-ß is the most potent naturally occurring inhibitor of cell growth. It exerts its action by binding to type I (TGFBR1) and type II (TGFBR2) transmembrane receptors located on the cell membrane.² Intracellular signaling is activated once TGF-ß has induced the formation of a TGFBR1/TGFBR2 heteromeric complex. TGFBR2 phosphorylates TGFBR1, resulting in activation of TGFBR1 kinase. Unrestricted cell growth due to a lack of growth inhibitory activity appears to be the most important of the possible consequences of a defect in TGF-ß function. This hypothesis was confirmed by the discovery that Tgfb1+/- and Tgfbr2+/- haploinsufficient mice resulting in decreased TGF-ß signaling have an increased susceptibility to develop cancer than their wild-type counterparts.^3,4 Specific targeting of the TGF-ß receptors in human cancer has been demonstrated by the identification of inactivating mutations in TGFBR2 in colon cancers and head and neck cancers, homozygous deletion of TGFBR1 in pancreatic and biliary carcinomas and in lymphoma, and a TGFBR1 tumor-specific mutation in breast cancer.^5–7 Additionally, restoration of functional receptor expression reverses the transformed phenotype of several human cancer cell lines that inherently lack functional TGF-ß receptors.^8,9 Clinical evidence that decreased TGF-ß signaling increases cancer susceptibility stems from studies showing that germline mutations of TGFBR2 and SMAD4 may predispose to the development of hereditary nonpolyposis colorectal cancer and juvenile polyposis, respectively.^10,11 Similarly, a polymorphism of TGF-ß1 resulting in increased TGF-ß signaling has been associated with a decreased risk of breast cancer.¹²

TGFBR1*6A is a polymorphic allele of the type I TGF-ß receptor (TGFBR1) that has a deletion of three alanines within a 9-alanine stretch of exon 1 coding sequence.¹³ Several epidemiological studies show an overrepresentation of both TGFBR1*6A heterozygotes and homozygotes among patients with a diagnosis of cancer.^14–18 Using mink lung epithelial cell lines devoid of endogenous TGFBR1, two groups of investigators have shown that TGFBR1*6A, compared to TGFBR1, is impaired as a mediator of TGF-ß antiproliferative signals.^14,15 Several recent studies have assessed the frequency of TGFBR1*6A among patients with a diagnosis of cancer and controls, with mixed results. Some studies confirmed this association; some did not.^18–21 The aim of this meta-analysis was to determine whether current epidemiological data support the hypothesis that TGFBR1*6A is a tumor susceptibility allele.

	PATIENTS AND METHODS

TOP ABSTRACT INTRODUCTION PATIENTS AND METHODS RESULTS DISCUSSION REFERENCES

 
Selection and Review of Studies
We performed a MEDLINE and CANCERLIT search from the database from January 1966 to April 2002. The search headings used were "Type I TGF-beta receptor," "TßR-I," "genetic polymorphism," "mutation analysis," "transforming growth factor-ß receptor," "mutation," "tumor susceptibility allele," "transforming growth factor," "TbR-I(6A)," "TbetaR-I(6A)," "TGFBR1*6A." The search was limited to human studies. A manual search of the bibliographies of all retrieved articles was performed by two independent investigators. All studies had to clearly describe the study population, including its ethnicity; if this was not clear, the corresponding authors were contacted and asked to provide the missing data. We excluded reviews, case series, and editorials, as well as studies assessing TGFBR1*6A in tumors. Overall, we identified seven studies investigating the TGFBR1*6A polymorphism in relation to cancer susceptibility. These studies are presented below. We were able to contact the senior author of each study except for one, and we received additional information that has been included in this report.

The study by Pasche et al,¹⁵ published in 1999, was the first case-control study to show an association of TGFBR1*6A {*6A} with cancer susceptibility.¹⁵ Subjects were from New York City and comprised 851 cases and 735 controls (Table 1). The patients and controls were similar with respect to sex, geographic location and ethnicity. Genotyping revealed that 88.9% of the controls carried the TGFBR1/TGFBR1 (*9A/*9A) genotype and10.6% the TGFBR1*6A/TGFBR1 (*6A/*9A) genotype, whereas no control subject carried the TGFBR1*6A/TGFBR1*6A (*6A/*6A) genotype. In the cancer group 84.1% of patients were *9A/*9A carriers and 14.6% *6A/*9A carriers. The *6A/*6A genotype was found in 1.1% of cases. The authors found a significantly greater proportion of *6A homozygotes and heterozygotes among cases than controls (Fig 1).

View larger version (18K): [in this window] [in a new window]   Fig 1. Meta-analysis of *6A for all cancers. The numbers under "Cases" and "Controls" represent *6A alleles out of all alleles.

Another case-control study by Pasche et al,¹⁵ performed on a Northern Italian population, consisted of 347 cancer patients and 50 controls. This case-control study, although a separate study, was presented in addition to the New York study in the Pasche et al article.¹⁵ The results showed a *9A/*9A frequency of 76% among controls and 84.7% among patients with a diagnosis of cancer. Furthermore, the *6A/*9A polymorphism was present in 24% of the controls and 14.7% of the cases. One patient with a diagnosis of breast cancer was homozygous for *6A.

In the study by Samowitz et al,¹⁹ published in 2001, 252 cases of colon cancer and 362 controls were studied in relation to the TGFBR1 polymorphism. Patients and controls from the state of Utah were similar by age, ethnicity, and sex (Table 1). The frequency of the *9A/*9A genotype in cases and controls was 80.2% and 81.5%, respectively. The heterozygous genotype (*6A/*9A) was present in 18.3% of cases and 16% of the controls, and the *6A/*6A genotype was found in 0.8% of cases and 1.4% of controls. Cases and controls had a similar proportion of *6A carriers. The investigators concluded that *6A was not associated with increased colon cancer risk.

The letter published by Stefanovska et al²⁰ in 2001 included 117 colon cancer patients and 200 controls. One hundred controls were newborn infants, and one hundred were elderly individuals. The study populations were from the Former Yugoslav Republic of Macedonia (FYROM) and were of similar ethnic status and the same sex. The frequency of the *9A/*9A genotype was 89.5% in the control population and 92.3% in the cancer cases. The *6A/*9A genotype was present in 10% of the controls and 6.8% of cases, and the *6A/*6A polymorphism was found in 1.0% of controls and 0.9% of cases. These results were not significant for any of the polymorphisms.

The study by van Tilborg et al²² published in 2001 was a case-control study, which included 183 normal blood donors and 146 patients with transitional carcinoma of the bladder, from the Netherlands. Patients were from similar ethnic groups but were not matched for age, sex, or ethnic status (Table 1). The investigators found no difference between cases and controls in relation to *6A/*9A (17.5% in controls v 17.1% in cases), whereas the *6A/*6A genotype was found in 3 (1.6%) controls and was absent from cases. The wild type *9A/*9A genotype was present in 80.9% of the controls and 82.9% of the cases. Therefore, there were no significant differences between bladder cancer cases and controls for the *6A polymorphism.

The final study included in our meta-analysis is from Baxter et al.¹⁸ This was a case-control study, which consisted of patients with breast (n = 355) and ovarian cancer (n = 304), endometriosis (n = 98), and 248 controls (Table 1). All cases and controls were of similar ethnic and geographic background. They were residents of Southampton and its surrounding suburbs, in the UK. The investigators found that 83.5% of controls carried the *9A/*9A genotype, compared to 76.5% of the cancer cases. The *6A/*9A genotype was present in 15.7% of the controls and 22% of the cases. The *6A/*6A genotype was present in 0.8% of controls and 1.5% of the cases. The authors found that *6A allelic frequency was significantly more common among patients with breast cancer than among controls.

Statistical Methods
We present the results of seven case-control studies as odds ratios (OR). If an association between the polymorphism and a specific diagnosis of cancer is positive, the OR will be greater than 1.0, and if the association is negative, the OR will be less than 1. The OR, their variance, and 95% confidence intervals (CI) were calculated using the methods of Mantel and Haenszel²³ and Robins et al.²⁴ The analysis was performed by pooling all studies together as well as according to specific tumor type. Because of incomplete data a statistical analysis by sex or ethnic background was not performed, but the available data are presented. Statistical analyses were done using SAS statistical software (SAS OnlineDoc, version 8; SAS Institute Inc, Cary, NC). Having demonstrated that there was a significant association between *6A homozygotes and *6A heterozygotes with cancer, we noticed that there was no association between *6A and bladder cancer. This led us to perform a secondary analysis excluding bladder cancer cases. Analyses were also done using the inverse variance method for combining OR. Results were similar, except for certain analyses of homozygotes, where cell frequencies were low.

Study Characteristics
The main features of the studies included in the meta-analysis are shown in Tables 1, 2, 3, and 4. The seven studies comprised 2,438 cases and 1,846 controls. All studies were case-control studies and were internally controlled for ethnicity. Taken together, the studies represent ethnically heterogeneous, but predominantly white populations.

	RESULTS

TOP ABSTRACT INTRODUCTION PATIENTS AND METHODS RESULTS DISCUSSION REFERENCES

 
Overall Data
Combined analysis of the seven studies showed that *6A allelic frequency was 28.8% higher among all cancer cases (0.0935) combined than among controls (0.0726; P < .001). *6A carriers had increased overall cancer risk (OR, 1.26; 95% CI, 1.07 to 1.49; Fig 1). Having demonstrated that *6A is associated with an overall increased risk of cancer, we proceeded with the analysis of each tumor group including at least 200 cases.

Breast Cancer
Three studies included patients with breast cancer^15,18 (Table 1). The total number of cases was 555. The overall OR for *6A carriers was 1.48 (95% CI, 1.11 to 1.96). *6A/*9A carriers had an OR of 1.45 (95% CI, 1.08 to 1.93). These findings indicate that *6A carrier status (due mainly to *6A heterozygotes) is significantly associated with an increased risk of breast cancer (Fig 2).

View larger version (11K): [in this window] [in a new window]   Fig 2. Meta-analysis of *6A for breast cancer. The numbers under "Cases" and "Controls" represent *6 alleles out of all alleles.

View larger version (10K): [in this window] [in a new window]   Fig 3. Meta-analysis of *6A for ovarian cancer. The numbers under "Cases" and "Controls" represent *6 alleles out of all alleles.

View larger version (13K): [in this window] [in a new window]   Fig 4. Meta-analysis of *6A for colorectal cancer. The numbers under "Cases" and "Controls" represent *6 alleles out of all alleles.

Hematologic Malignancies
Only one study investigated the possible association between *6A and hematologic malignancies.¹⁵ It comprised 219 patients, 80 with non-Hodgkin’s lymphoma, 26 with acute lymphocytic leukemia, 34 with acute myelogenous leukemia, 6 with chronic lymphocytic leukemia, 22 with chronic myelogenous leukemia, 18 with Hodgkin’s disease, and 33 with multiple myeloma. Of those, one patient (0.47%) with non-Hodgkin’s lymphoma was homozygous for the *6A allele and 36 patients (16.4%) were heterozygous for the *6A allele. The control group had a 10.6% frequency of the *6A/*9A genotype, and no control patient was homozygous for the *6A allele. The OR for the *6A/*9A genotype was 1.60 (95% CI, 1.06 to 2.41). There was no difference in *6A allelic frequency among the various subtypes of hematologic malignancies (P = .10). This limited study of pooled patients with various forms of hematological malignancies suggests that *6A may possibly be associated with some hematologic malignancies.

Analysis Excluding Bladder Cancer Cases
Having demonstrated that *6A was not associated with bladder cancer, we performed a meta-analysis that excluded bladder cancer cases to more accurately determine the magnitude of the cancer risk associated with *6A. We found that *6A homozygotes’ risk for cancer was twice that of *6A heterozygotes (OR, 2.53; 95% CI, 1.39 to 4.61 v OR, 1.26; 95% CI, 1.04 to 1.51).

Age Distribution
Due to insufficient data, we were unable to perform an analysis based on age distribution. Cases and controls from the Samowitz et al study had similar age distribution.¹⁹ There was no information available for the studies by van Tilborg et al (2001) and Pasche et al (1999 Italian study). In the study by Stefanovska et al (2001),²⁰ the mean and median ages of the controls were 37.8 and 24 years, respectively, and those of the patients were 63.5 and 65 years, respectively. In the study by Pasche et al (1999 US study), the mean ages for cases and controls were 56 and 35 years, respectively.¹⁵ In the study by Baxter et al, the mean age of the ovarian cancer cases was 62 years (range, 23 to 92 years); the mean age of breast cancer cases was 38 years (range, 19 to 79 years); and the mean age of controls was 39 years (range, 18 to 84 years). Overall, the mean age of cases was higher than that of controls. If *6A acts as a tumor susceptibility allele, cancer incidence among *6A carriers is likely to increase more rapidly with age than among noncarriers. Thus, the younger mean age of controls could result in a bias toward the null hypothesis, resulting in an even stronger association than observed in this meta-analysis.

Sex Distribution
The studies by Baxter et al and Chen et al were matched by sex, given that they included only female cases and controls; the studies by Samowitz et al and Pasche et al (1999), as well as Stefanovska et al, which represent 1,945 cases and 1,586 controls, were controlled for sex. There was no sex information in the Italian study by Pasche et al (1999), nor for the study by van Tilborg et al.

Ethnic Distribution
Studies included in our meta-analysis included various ethnic groups— predominantly white patients and, to a minor extent, black patients. We were not able to obtain any ethnic data on the van Tilborg et al study²² and the Pasche et al 1999 Italian study.¹⁵ The studies by Baxter et al, Stefanovska et al, and Samowitz et al included essentially only whites. The study by Chen et al included two subpopulations.¹⁴ One was the Jamaican population, which included only black females; the other was a study conducted in six East Coast centers in the US. This study included subjects of different ethnic backgrounds, but further details could not be obtained.

Two BRCA1, one BRCA2, and one APC mutations are found predominantly among Ashkenazi (Eastern European) Jews and only rarely in the remainder of the white population. For this reason, we analyzed our subjects according to whether they were or not of Jewish decent. For some studies, we did not have any available information. In the US study by Pasche et al (1999), data were available for both cases and controls. From the 617 cases for which data were available, 166 were Jews. There were 146 *9A/*9A (88.0%), 17 *6A/*9A (10.2%), and 2 *6A/*6A (1.2%). Of the 605 non-Jews, 503 were genotyped as *9A/*9A (83.1%), 98 *6A/*9A (16.2%), and 4 *6A/*6A (0.03%). An analysis of these data did not show any difference in the frequency of *6A between Jews and non-Jews (P = .09).

	DISCUSSION

TOP ABSTRACT INTRODUCTION PATIENTS AND METHODS RESULTS DISCUSSION REFERENCES

 
This meta-analysis of seven case-control studies shows that *6A correlates significantly with breast and ovarian cancer as well as with colon cancer in the US only (Table 4). Individually, only two of the seven studies included in our meta-analysis concluded that there was an association between *6A and cancer. Although this argues against a publication bias of positive studies only, we conducted a comprehensive search to identify unpublished studies. Publication bias is unlikely in the current meta-analysis since all studies that have assessed the relation of *6A with cancer were included. A funnel plot of any *6A mutation for all cancers (Fig 5) supports the lack of publication bias in that studies are evenly distributed in each half of the plot. Small negative studies with OR that are relatively small in magnitude are included in this analysis.

View larger version (11K): [in this window] [in a new window]   Fig 5. Funnel plot of the log odds ratio (horizontal axis) versus precision (vertical axis). Each circle represents one study. Larger circles indicate larger sample sizes.

We found a significant association of *6A with cancer. After exclusion of all patients with bladder cancer, a tumor type unequivocally not associated with *6A, we were able to better estimate the OR for cancer associated with *6A carrier status. The OR for *6A homozygotes was twice that of *6A heterozygotes suggesting an allelic dosing effect. The data indicate that *6A results in a low-penetrance syndrome for heterozygotes and a low-frequency, low-to-moderate penetrance syndrome for homozygotes. Our combined results reveal that more than one in eight healthy individuals and one in six patients with cancer is a *6A carrier. This sets a new paradigm and establishes *6A as the first high-frequency, low-penetrance tumor susceptibility allele affecting a variety of ethnic groups. Our US results suggest that 8,099 (5.5%) of 148,300 new colon cancer cases per year may be attributable to *6A. Our combined results indicate that 14,555 (7.1%) of 205,000 and 2,515 (10.8%) of 23,200 new US cases of breast and ovarian cancer per year may be attributable to *6A. Our findings support the growing evidence that alterations in TGF-ß signaling as a result of polymorphisms or variants may affect risk for certain forms of cancer.^10–12,25 This study warrants additional large case-control studies as well as studies among families with increased incidence of breast, ovarian and colon cancer.

	ACKNOWLEDGMENTS

 
This work was supported by grants CA76156–04, CA082516–01A2, and CA89018 from the National Cancer Institute (B.P.), an Avon Breast Cancer Career Development Award (B.P.), a grant from the Mander Foundation to the Cancer Genetics Program, and a grant from the Lymphoma Foundation (K.O.).

	REFERENCES

TOP ABSTRACT INTRODUCTION PATIENTS AND METHODS RESULTS DISCUSSION REFERENCES

 
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Submitted November 22, 2002; accepted June 6, 2003.

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