Congenital talipes equinovarus (CTEV) is a common developmental disorder of the foot, affecting between 1 and 4.5 per 1000 live births. The aetiology is not well elucidated. While both genetic and environmental factors are implicated, no specific genes have been identified and little is known about environmental risk factors.We conducted a case-control study of idiopathic congenital talipes equinovarus (ICTEV) in the United Kingdom. 194 cases and 60 controls were recruited. Pedigrees were obtained for 167 cases.The rank of the index pregnancy, maternal education and caesarean delivery were significantly associated with ICTEV risk in a multivariate model. There were suggestions that maternal use of folic acid supplements in the three months before the pregnancy decreased ICTEV risk, and that parental smoking during the pregnancy increased risk, although the associations were not statistically significant. One quarter of pedigrees showed a family history of CTEV, and autosomal dominant inheritance was suggested in some of these.Uterine restriction did not appear to have a strong influence on ICTEV development in our study. Large population-based studies are needed to clarify the aetiology of this common developmental disorder. Congenital talipes equinovarus (CTEV) is a common developmental disorder of the foot. It is defined as fixation of the foot in cavus, adductus, varus and equinus  with concomitant soft tissue abnormalities . CTEV maThe UK Talipes Study was designed to investigate the aetiology and describe patterns of management of ICTEV in the UK. In this article we report the epidemiological findings and describe pedigree analyses of families affected by ICTEV.st July 1993 and 1st July 1997. All children had been diagnosed with ICTEV by a treating orthopaedic surgeon. The source of the first series was the Scottish Talipes Register, which was established in 1993 and included all hospitals involved in CTEV management [Three groups of children with ICTEV, and their parents, were recruited between 1nagement . The souFamilies were approached by letter and invited to participate by attending their local hospital for orthopaedic assessment and interview and by completing questionnaires. The orthopaedic assessment comprised notes review and clinical examination in order to confirm non-syndromic status. Family history was recorded on a pedigree for each family that attended for interview. Each family was approached only once. The questionnaires included sections on the nature of the ICTEV , pregnancy and birth of the affected (index) child, maternal reproductive history, parental smoking and alcohol use, parental education and ethnicity. Families were given the option of completing the questionnaires without attending interview. We recruited unaffected cousins as controls. These were sex and age-matched to cases where possible. The questionnaires were sent by the index family to the potential control family, and returned directly to the investigators.Analyses were conducted using Stata 8 . We carrThe study was approved by the Grampian Research Ethics Committee.458 ICTEV subjects were eligible for the study and 194 participated by completing the questionnaires (response rate 42%); 167 also attended for interview. Cousin controls were recruited for 60 (31%) participants. The characteristics of the cases and controls are summarized in Table The ratio of males to females was 2.1:1. Fifty one percent had bilateral ICTEV . Of the unilateral cases, more had right foot involvement than left . Females who were affected unilaterally were more than twice as likely to be affected on the right than the left, whereas in males left and right sides were equally affected ; the differences were not statistically significant. Three quarters of cases were born in 1990 or later.There were ten twin pregnancies amongst the cases (5%), nine of which were reported to be non-identical; the zygosity was unknown for the tenth. None of the non-index twins had CTEV. There were no multiple pregnancies amongst controls. Eighty one percent of ICTEV deliveries were vaginal with cephalic presentation, 2% were vaginal with breech presentation and 17% were caesarean. In controls, the corresponding figures were 93%, 2% and 5%. Odds ratios, adjusted for sex and year of birth, were significantly raised for caesarean section  of case and 25% (15/59) of control mothers had had at least one miscarriage; 3% (5/191) of case and 2% (1/60) of control mothers had had a stillbirth and 17% (33/192) of case and 15% (9/60) of control mothers had had at least one abortion.The median age of mothers at the birth of the index child was 28 years for both cases and controls . The median age of fathers was 31 years for cases (range 17–58) and 30 years for controls (range 19–50). The median age of the mother at the birth of her first child was 24 years for cases (range 15–38) and 25 years for controls (range 16–38).The index pregnancy was the first for 42% of case and 53% of control mothers. The risk of ICTEV increased significantly with the rank of the pregnancy (p for trend = 0.03). Adjusting for maternal age made minimal difference to the risk estimates. Further adjustment in the final multivariate model (see below) increased the risk estimates slightly. The mean parity was 2.2 for case (range 1–6) and 2.1 for control (range 1–6) mothers.Around one quarter of mothers  reported using nutritional supplements containing folic acid in the three months before the index pregnancy or in the first trimester. Use increased with the year of birth of the child and was 5% for births 1963–1979, 13% for 1980–1989, 23% for 1990–1994 and 28% for 1995–1999. The increase over time was statistically significant (p for trend <0.001). Use of these supplements was associated with a modest, but non-significant, reduced risk of ICTEV. When the analysis was restricted to folic acid taken in the three months prior to the pregnancy the effect was more pronounced , but remained non-significant.More case than control parents reported smoking during the index pregnancy ; the differences were not significant. Whilst the association with maternal smoking was stronger in female than male children , the risk estimates were not statistically significant, nor was a formal test of interaction between sex and maternal smoking significant (p = 0.41). Fewer case than control mothers reported drinking alcohol during the index pregnancy , but the differences were not significant.More case than control mothers had further education or training , although the difference was not statistically significant. Fathers of cases and controls had similar levels of further education.Table 21 7.53, p = 0.01) but the test for interaction between sex and folic acid use was not significant (p = 0.18). Folic acid use was not associated with laterality.In the case-only analysis there were no statistically significant associations between sex and laterality , nor between the sex of the case and multiple birth, type of birth, rank of the pregnancy, parental smoking, maternal alcohol use or maternal education. In addition, there were no associations between these variables and laterality . For maternal use of folic acid, there were significantly fewer females than expected in the group whose mothers used supplements  a 1st degree family history and 4.2% (7/167) a 2nd degree family history; 10.8% (18/167) had a 1st or 2nd degree family history, and 1.2% (n = 2) a 1st and 2nd degree family history. Of the 13 families with a 1st degree family history, there were six child-parent pairs, six sib-pairs and one trio in which both a parent and a sib were affected. Six families had three or more cases of CTEV in the family (3.6%); the pedigrees of three of these suggested autosomal dominant inheritance whilst the mode of inheritance was uncertain in the remainder.A pedigree was available for 167 families. Families with and without pedigrees were compared for the variables in Table st–3rd degree family history. Eight percent of males (n = 9) and 7.1% of females (n = 4) had a 1st degree family history, and 5.4% of males (n = 6) and 1.8% of females (n = 1) had a 2nd degree family history. None of these differences reached statistical significance (data not shown).Male index cases were more likely to have a family history than female index cases; 18.9% of males (n = 20) and 12.5% of females (n = 7) had a 1st degree relatives  and 0.5% of 2nd degree relatives  were reported to have CTEV. The absolute risk of any 1st or 2nd degree relative being affected was 1.1% . Slightly more relatives of male  than female  index cases had ICTEV. The risk to each male relative, regardless of the sex of the case, was 1.2%  and for each female relative was 0.9% . When the case was male the risk of CTEV was higher for male  than female  relatives. When the index case was female the risk of CTEV was slightly higher for female relatives . None of these differences reached statistical significance (data not shown).Of the 167 index cases, 2.7% of 1st–3rd degree family history and the variables sex, laterality, maternal alcohol and maternal smoking showed no associations. There was a significant association between family history and paternal smoking , but not with maternal smoking .Cross-tabulations between 1The sex ratio in our study and the predominance of bilateral disease mirrors the pattern seen in other studies of children with ICTEV -19. Of t2 19.99, p<0.001). If this rate held amongst controls, we would have expected one twin birth. Uterine restriction is the earliest known hypothesis proposed to account for CTEV, with references dating back to Hippocrates in the 5th century BC. Our finding of a higher twin birth rate in cases than controls could be considered to support this hypothesis. However, we also found a positive association between risk of ICTEV and the rank of the index pregnancy; if it is assumed that the primagravid uterus is more restrictive than a multigravid uterus, uterine restriction is not a major factor in our series.Five percent of cases but no controls were twin births. The rate of twinning in our cases was significantly higher than the Scottish rate for 1996–2000 of 1.4% [The rate of caesarean delivery was significantly higher in our cases than controls (17% vs. 5%). We did not have information on why the caesareans were performed. Twin births accounted for some of the excess, as 6/10 twin births were delivered by caesarean section, but the association remained when the analysis was restricted to singletons. The frequency of caesarean deliveries for singleton ICTEV pregnancies in our study (14%) is consistent with figures for the Scottish population, taking into account the range of years of birth covered by our study . Thus thMaternal education and birth rank were also significant in our final model, both being positively associated with the risk of ICTEV. Maternal education, which may be seen as an indicator of socioeconomic status, has been infrequently reported and findings are inconsistent with positive , inverseFolate appears to play a role in some congenital malformations, particularly neural tube defects . An assoMaternal smoking has been implicated as a risk factor for CTEV ,28-30. Ist–2nd degree relatives as this was considered the most reliable data.One quarter of children had a family history of CTEV, which corresponds with the upper limit of the estimates reported in two series in the USA ,32, but st degree family history differs widely. Our frequency of 7.8% is similar to that of 7.2% reported by Honein et al [st degree relatives in our study was 2.7%, within the range reported in the literature (2.1–3.6%) [The proportion of CTEV families reported to have a 1in et al , whilst in et al  and Wynnin et al . The abs.1–3.6%) ,16,30,33We found an increased risk of a family history of CTEV in male compared with female index cases, although the difference was not statistically significant. The highest risk of CTEV in our study was in male relatives of male probands, contrary to Wynne-Davis (1964) who found the highest risk in male relatives of female probands in a series of 635 UK cases . We had The heterogeneity between studies in reported risk factors for CTEV may be due to many factors including definitions of CTEV, whether non-idiopathic cases are included, differences in study design, and chance as some series are relatively small. Participation bias may also play a role, possibly with multi-case families or those with more severely affected children more likely to take part. Studies of family history of CTEV are likely to vary in the degree of evidence considered sufficient for a 'diagnosis' of CTEV in relatives. The aetiology of CTEV is undoubtedly complex and it is probable that both environmental and genetic factors are involved. The genetic influence itself is likely to be complex and may involve multiple genes or genetic pathways, themselves interacting with environmental factors.We have given an epidemiological picture of a large series of children with ICTEV in the UK. Autosomal dominant inheritance seems to be present in some of the families, although the mode of inheritance was unclear for the majority of multi-case families. Cesarean delivery, birth rank and maternal education were all significantly associated with the risk of ICTEV. Uterine restriction does not seem to play a major role in our series. We found significant associations between family history and paternal smoking, which may indicate an interaction between smoking and an as-yet unknown gene or genes. Studies of unbiased series from different geographical and ethnic groups are essential for elucidating both the genetic and epidemiological factors involved in the aetiology of this common development disorder.CTEV – congenital talipes equinovarusICTEV – idiopathic congenital talipes equinovarusOR – odds ratio95%CI – 95% confidence intervalThe author(s) declare that they have no competing interests.AC conducted the analysis and drafted the paper. SB and DC conducted orthopaedic assessment of participants, collected the data and commented on the paper. LS contributed to study design, running the study, advised on analysis and contributed to paper writing. NM conceived the original study and commented on the paper. ZM contributed to study design, running the study and to paper writing. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


A truncating variant, 1100delC, in check point-kinase CHEK2, has been identified as a risk factor for familial and sporadic breast cancer. The prevalence in healthy non-breast cancer cases is low and varies between populations.CHEK2 1100delC in 763 breast cancer patients with a defined family history and 760 controls from the Stockholm region. The breast cancer patients originated from; a population-based cohort (n = 452) and from a familial cancer clinic (n = 311), the detailed family history was known in both groups.We analyzed the prevalence of The variant was found in 2.9% of the familial cases from the population-based cohort and in 1.9% from the familial cancer clinic. In total 2.2% of the patients with a family history of breast cancer carried the variant compared to 0.7% of the controls (p = 0.03). There was no increased prevalence in sporadic patients (0.3%). The variant was most frequent in young familial patients . The mean age at diagnosis of variant carriers was 12 years lower than in non-carriers (p = 0.001).CHEK2 1100delC exists in the Swedish population. The prevalence is increased in familial breast cancer and the variant seems to influence age at onset.In conclusion,  BRCA1, BRCA2, p53, ATM and PTEN account for less than 25% of the familial risk for breast cancer while the remainder are still genetically unexplained despite large efforts in research [BRCA1/2 aggregation of breast cancer, and the effect of low penetrant genes may also at least partly explain sporadic breast cancer [CHEK2 (1100delC), a G2 checkpoint-kinase, that is involved in cell cycle control and DNA-repair, was identified as a risk factor for breast cancer in two independent studies using this approach [BRCA1/2 cases and healthy controls, and both studies were highly significant despite the fact that the variant is rare  and confers a relative risk of around 2 [Apart from gender, family history is the most important risk factor for breast cancer. Mutations in the known high-risk genes t cancer . Associaapproach -8. The varound 2 ,8. More around 2 ,8. The varound 2 -9. A sub [CHEK2 10delC, a and a family history of breast cancer or had been collected as part of previous research on familial breast cancer [311 independent familial cases collected at the Department of Clinical Genetics at Karolinska Hospital, Stockholm were used for the study. These patients had either been referred due to a breast cancer diagnosis t cancer ,13.For the Familial Risk cohort only age at diagnosis was available (missing in 16 patients).BRCA1 or BRCA2 mutations were excluded from the study.Patients with a surgically treated primary invasive breast cancer admitted to the Department of Oncology at Huddinge Hospital and Söder Hospital (covering the population of southern Stockholm of 850 000 people) from October 1998 to May 2000 were asked participate in a study on genetic risk factors from breast cancer . Family st or 2nd-degree relatives with breast cancer in addition to the proband, regardless of age, were classified as familial breast cancer. In the Familial risk cohort 168 cases had one 1st or 2nd degree relative, while the remainder had more than one 1st or 2nd degree relative. In the population-based cohort 104 cases had one relative and 35 more than one relative with the disease.Cases with at least one 1The mean age at diagnosis in the Familial risk cohort was 54 years (24–92 years). In the Population-based cohort the mean age was 60 years (27–88 years) and there was no statistically significant difference between familial and sporadic cases (59 and 61 years respectively).As controls we used DNA from 760 geographically matched blood-donors of mixed gender collected as a control material for association studies at Karolinska University Hospital, Stockholm, Sweden.The Ethical Committee at the Karolinska Institute approved the study.CHEK2 gene (1100delC) was performed on controls and patients by PCR with a primer set that is specific for the mutation , 125 μM of each dNTP, 4 pmol of each primer, 0.5 units of Amplicon III Taq Polymerase , and 25 – 50 ng of template DNA. The PCR conditions were: initial denaturation at 95°C for 5 min. hereafter 40 cycles at 95°C for 20 sec, 59°C for 20 sec and 72°C for 20 sec followed by 5 min. of extension at 72°C.Genotyping of the most common mutation in the The two reactions were run separately each in multiplex with a control PCR (Control-F: 5'-gtc aaa gcc acc agt tac agt-3' and Control-R: 5'-ttc ccc acc act tta ctg ac-3') resulting in a product of 309 base pairs on chromosome 4. The products were separated on 2.1% SeaKem LE Agarose  gels. In order to examine for homozygosity for the variant, PCR for wild type allele and control were performed in cases with the CHEK2 1100CdelC variant.Fisher's exact test was used to compare categorical data. Continuous data were compared using the two-sample t-test.CHEK2 1100delC was 1.9% in the familial risk Cohort and 1.1% in the population-based cohort compared to 0.7% in the controls (p = 0.09 and 0.51 respectively). 2.2% of all the familial cases carried the variant (p = 0.03), corresponding to an odds ratio of 3.4 (95% CI 1.2–10.1).  and 10 (population-based cohort) years lower in carriers than in non-carriers; 42 (24–55) vs. 54 years in familial risk cohort (p = 0.01) and 50 years (38–65) versus 60 years in the population-based cohort (p = 0.06). If the groups were combined the difference in mean age at onset was 12 years .In three families of carriers from the Familial Risk cohort, DNA was available for more family members. In two of these families, the mutation carrier status did not segregate with disease Figure .The five heterozygotes in the Population-based cohort all had unilateral ductal breast cancer without recurrence or secondary malignancies with 6 years follow-up.Only one CHEK2 1100delC carrier did not have a family history of breast cancer Table .We have evaluated the prevalence of the CHEK2 1100delC variant in two cohorts of breast cancer patients from the Stockholm region, one familial and one population-based, both with a well-defined family history and in controls. In our study CHEK2 1100delC was associated with familial breast cancer  and confirm previous results of the variant as a risk factor for breast cancer.BRCA1/2 families also harboring colon cancer cases (18%) and in some highly selected high-risk breast cancer families (9–11%) [In the original studies from Finland and the Netherlands the prevalence in familial breast cancer was 3.1–5.5 % ,8. The hfamilies –11% [9,1 (9–11%) ,16. The  (9–11%) -20.CHEK2 1100delC among sporadic cases, the cohort however too small to draw any firm conclusions. The difference in prevalence in our two cohorts regarding familial breast cancer (1.9% vs. 2.9%) is also based on very few cases and we refrain from analyzing this difference.Sporadic breast cancer cases have been reported to have a higher prevalence of the variant than healthy controls as demonstrated in the large pooled study by the Breast Cancer Consortium  . In our CHEK2 1100delC in our controls was 0.7%, which is consistent with previous studies in the Swedish population (0.4–1%) and the large pooled control material of more than 9000 controls, mainly of Western European origin (0.7%) [The prevalence of n (0.7%) -23.CHEK2 1100delC carriers were markedly younger at diagnosis compared to non-carriers, even though the difference was of borderline significance in the two groups of breast cancer cases and the material small. Several previous studies support this finding including the large pooled analysis where mutation prevalence decreased with increasing age at diagnosis [In our material the iagnosis ,24-26. Oiagnosis ,16,27,28iagnosis .CHEK2 was originally suggested to be a high-risk gene for the Li Fraumeni syndrome [syndrome . There isyndrome ,16,30,31Our results, generated in a material with a well-defined family history including the paternal side, support the role of a modifier, as there was no accumulation of the variant in true sporadic cases, and the variant seemed to influence the age at diagnosis in carriers. Our material is however relatively small, and this conclusion would need to be verified in a larger material.In conclusion the CHEK2 1100delC variant was significantly more frequent in familial cases assumed to modify an underlying hereditary fault Since the effect of the variant is modest and the variant rare, there is no need for CHEK2 1100delC screening at present but the variant might prove interesting in combination with other genetic/non-genetic factors in the future.The author(s) declare that they have no competing interests.SM collected the population-based material, participated in the design of the study, performed the statistical analysis and wrote the manuscript.HE participated in development of the method and in the molecular genetic analysis.AL conceived the idea of the study, collected the familial risk cohort material and participated in the analysis of the study.MLB participated in the design of the study, performed the molecular genetic analysis, and participated in the analysis of the results and in drafting the manuscript.All authors read and approved the final manuscriptThe pre-publication history for this paper can be accessed here:


The relation of breast cancer to social and dietary variables was evaluated in a case-control study of 368 women with breast cancer admitted to the General Hospital of Pordenone  and 373 age-matched controls. Occupation was related to the risk of breast cancer, housewives and non-manual workers  showing relative risks of 1.7 and 2.4 respectively when compared to women occupied in agriculture. The role of education was apparently less important, and not statistically significant. The risk was higher in women who were obese, the trend of increasing risk with increasing body mass index being confined to post-menopausal women. When indicators of dietary fat intake were analysed, a significantly increased risk was found with more frequent consumption of milk and dairy products but the risk estimates were only slightly above unity with reference to meat consumption. Women who drank alcoholic beverages showed a relative risk of 2.5 compared to women who had never drunk, when allowance was made for all identified potential confounding factors. The association between alcohol and breast cancer was not explained by the other dietary variables considered, and the risk estimates were higher for women who drank more wine, or more than one type of alcoholic beverage. Thus, the findings of the present study give evidence in favour of the hypothesis that alcohol consumption is related to the risk of breast cancer.


Although impacts upon gastric cancer incidence of race, age, sex, and Lauren type have been individually explored, neither their importance when evaluated together nor the presence or absence of interactions among them have not been fully described.This study, derived from SEER  Program of the National Cancer Institute) data, analyzed the incidences of gastric cancer between the years 1992–2001. There were 7882 patients who had developed gastric cancer. The total denominator population was 145,155, 669 persons . Patients with multiple tumors were evaluated as per the default of the SEER*Stat program. 160 age-, five year period (1992–1996 vs 1997–2001)-, sex-, race (Asian vs non-Asian)-, Lauren type- specific incidences were derived to form the stratified sample evaluated by linear regression. (160 groups = 2 five year periods × 2 race groups × 2 sexes × 2 Lauren types × 10 age groups.) Linear regression was used to analyze the importance of each of these explanatory variables and to see if there were interactions among the explanatory variables.Race, sex, age group, and Lauren type were found to be important explanatory variables, as were interactions between Lauren type and each of the other important explanatory variables. In the final model, the contribution of each explanatory variable was highly statistically significant . The regression equation for Lauren type 1 had different coefficients for the explanatory variables Race, Sex, and Age, than did the regression equation for Lauren type 2.The change of the incidence of stomach cancer with respect to age for Lauren type 1 stomach cancer differs from that for Lauren type 2 stomach cancers. The relationships between age and Lauren type do not differ across gender or race. The results support the notion that Lauren type 1 and Lauren type 2 gastric cancers have different etiologies and different patterns of progression from pre-cancer to cancer. The results should be validated by evaluation of other databases. Worldwide, the stomach is the second most common site of origin of cancer . AlthougThe Surveillance, Epidemiology, and End Results (SEER) Program of the National Cancer Institute is an authoritative source of information on cancer incidence and survival in the United States that currently collects and publishes cancer incidence and survival data from 14 population-based cancer registries and three supplemental registries covering approximately 26 percent of the US population; the SEER website provides extensive information about it .The study used SEER to evaluate the contributions of age, sex, race (Asian vs non-Asian), year of diagnosis (1992–1996 vs 1997–2001), and Lauren type to gastric cancer incidence. The study showed Lauren type 1 tumor incidence increased with respect to age in a different way than did Lauren type 2 tumor incidence; the regression equations that described these relations were the same for men and women and for Asians and non-Asians. Incidence was considered in terms of the natural logarithms of the rates of development, over a five year period, of stomach cancer.The SEER data base, SEER 11 Regs + AK Public-Use, Nov 2003 Sub for Expanded Races (1992–2001) was used . The ana-5. From 1997–2001, of 366,766 Asian women aged 65–69, 55 developed Lauren type 1 gastric cancer: the rate was 1.60 × 10-4.Hence, SEER generated 160 numbers pairs. Each pair comprised a number of persons who developed stomach cancer and a number of persons in the denominator. For each set a rate was calculated by dividing the number of persons with stomach cancer by the number of persons in the denominator. For example, from 1992–1996, 35 Lauren type 1 stomach cancers were observed among 2,752,873 non-Asian men ages 40–44: the rate was 1.27 × 10R was used for data analysis.Counts and population provided by SEER*STAT were used to calculate rates. Preliminary data analysis showed the distribution of rates lacked a normal distribution. One cell had no persons with cancer; to take the natural logarithms of rates in a circumstance in which a zero cell is present, one may increase for all cells the numerator and the denominator by 0.5 :ln(ca) = ln [(persons with cancer + 0.5)/(persons in denominator + 0.5)]P < 0.05. Neither higher-order explanatory variable interactions nor non-linear relationships of explanatory variables to the response variable were considered in this analysis.All models tested were linear regression models with the response variable being the logarithms of the cancer rates, as defined above. Analysis of variance determined which model best reflected the data. Covariates were always added to regression; the ratio of the change in the residual sum of squares (ΔRSS) to the RSS before the covariate was added was compared with an F test. Independent (explanatory or predictor) variables included: A) Five year period , B) Sex , C) Race , D) Lauren type , E) Age Group . All ten possible two-way variable interactions were assessed. The null hypothesis was rejected if z > 600, P < 1 × 10-10. A residual plot showed the model lacked a good fit.Evaluation of the data precluded the performance of Poisson regression: the mean number of patients with cancer was 49; the variance was 2224. When Poisson regression was tried with population as an offset, with or without the zero cell, all five potential explanatory variables  and all ten potential first order interactions were associated with the counts; each explanatory variable had an associated Standardized residuals were calculated by R. To test the assumption that the standardized residuals were normally distributed, a Shapiro-Wilks test was performed. To test the assumption that the mean of the standardized residuals was 0, a t-test was performed. To test the assumption that the standardized residuals had constant variance with respect to the fitted values, the standardized residuals were divided into quartiles and Bartlett's test for homogeneity of variances was performed. A data point was considered an outlier if its studentized residual, calculated by R, was greater than 3; dffits (a measure which gives greater weight to outlying observations) and Cook's distance (a measure of impact of the respective case on the regression equation), calculated by R were used to assess leverage; outliers that are found to have high leverage, large dffits and/or Cook's distances, are considered bad leverage points and are removed from the analysis.The distribution of persons who developed cancer, stratified by Lauren type, five year period, sex, and race, is displayed in Table Initial evaluation showed rates lacked a normal distribution . As discussed in the model adequacy section, the use of the logarithms of the rates yielded a model that fulfilled the assumptions of linear regression, once an outlier was removed; the residuals of that model did not prove to lack a normal distribution, did not have a mean that proved to differ from zero, and did not prove to lack homogeneity of variance. Table The final model did not include five year period as an explanatory variable because 1) there was no association between the natural logarithms of the cancer rates and the five year period and 2) there was no demonstrated interaction between five year period and any other explanatory variable. The final model, displayed in Table Two regression equations, using the values for the final model without the outlier, express the results:Lauren type 1ln(ca) = -14.15 - 1.07 × sex + 1.74 × race + 0.52 × age groupLauren type 2ln(ca) = -11.55 - 0.23 × sex + 0.97 × race + 0.28 × age groupFor the above equations:• ln(ca) is the response variable, the natural logarithm of the stomach cancer rates.risk factors)• Sex, Race, and Age are explanatory variables , then, for Lauren type 1 stomach cancer, mean ln(ca) increases by about 1.74 if the group is Asian; for Lauren type 2 stomach cancer, the corresponding increase is 0.97.• The numbers in front of the explanatory variables are called regression Figures There was only one outlier, a data point with a studentized residual over 3. No non-Asian women, aged 45–49 years, developed Lauren type 1 stomach cancer; it was the only zero cell. The studentized residual, -5.40, corresponded with the largest Cook's distance, 0.183, and the dffit, -1.316, with the largest absolute value. The outlier was a bad leverage point. Although results for the outlier are displayed in the tables, the high leverage meant that the outlier should be excluded from the final analysis. Removing the outlier yielded the same choice of covariates.Quantitative assessments of model adequacy are displayed in Table This study found, in the SEER database, that Lauren type 1 and Lauren type 2 stomach cancers differ to such a degree that different regression equations are required to explain variations in their incidences. Sex, race (Asian or non-Asian), and age are explanatory variables, but the equations that relate these explanatory variables to the incidence of each Lauren type differ. Recent epidemiologic studies well support the rationale of the current study, namely to evaluate year of diagnosis (in this case five year period), sex, race, age and Lauren type. The articles also support the need for evaluation of interactions and also provide interesting thoughts about the limitations of administrative databases and other factors that should be considered in future studies.Boyle  found stMarmo , TurkdogCiliated metaplasia, a precursor to stomach cancer, occurs at different rates in the Pacific and Atlantic basins . Yao 2727 showedAs to black race, some have suggested that Caucasians are more likely than blacks to develop gastric cancer that arises in the cardia and that blacks are more likely than Caucasians to develop gastric cancer that arises outside the cardia ,33. TherOlder persons more likely develop ciliated metaplasia than do young persons . MultiplLoss of CDX2 may represent a marker of tumor progression in early gastric cancer and carcinomas with an intestinal, but not a non-intestinal phenotype . The freThe above discourse allows one to appreciate the limitations and utility of the study. SEER is, like many of the sources of the other studies, an administrative database. Administrative databases lack a review of histopathology; the added loss of precision is unavoidable because such a review would increase the expense of any such study and decrease participation by hospitals, largely invalidating its results. As expected, specific program codes are not available on line for investigators, reviewers, readers, and editors to explore issues that may be important to them, such as the means of creation of a denominator in rate calculations. The website is excellent, but might also include readily accessible links to the data registries themselves and their policies and procedures, so investigators, reviewers, editors, and readers can satisfy any questions they might have as to such matters as data collection or the particular manner of dealing with multiple primaries for a particular study. No administrative database can keep a record of such things as H pylori rates, genetic markers, food intake, or any of the other above miscellaneous factors identified. As with any study, the number of factors that can be evaluated is limited both for reasons of data collection and for statistical reasons having to do with sample size; for this reason, a global explanation encompassing all potential factors cannot be expected. The most any epidemiologic study can offer is a partial explanation of complex phenomena. Most vital, the above referenced recent studies show that any conclusion derived by examination of a particular population must be verified by evaluation of multiple populations. This is because factors that are important in one population may be unimportant in another population; only by repeating an analysis in multiple populations can an epidemiologic conclusion be considered verified. Notwithstanding these caveats, such studies of epidemiology have great practical significance; Marmo  used sucIn summary, two regression equations were derived from the SEER database to explain differences in stomach cancer incidence, one for Lauren type 1 stomach cancers, one for Lauren type 2 stomach cancers. Each regression equation revealed a simple relationship between the natural logarithm of stomach cancer incidence rates and age. These equations were the same for men and women and for Asians and non-Asians. These results should be verified by similar evaluations conducted in other populations.The author(s) declare that they have no competing interests.MW and YZ performed the regression analysis. MC, MW, and EF conceived of the study, and participated in its design and coordination and helped to draft the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


The search for biologic endpoints and biomarkers in the study of breast cancer risk assessment and risk reduction strategies has led to an interest in obtaining cytologic information and other biomarkers from nipple aspirate fluid (NAF).This descriptive study examined factors associated with an increased ability to obtain NAF in a cohort of 3043 women between the ages of 15 and 89 years of age. The majority of women were between the ages of 30–49 (N = 1529/50.2%). Variables examined in relation to obtaining fluid include: age, marital status, age at menarche, menopausal status, a history of pregnancy, a history of breast-feeding, estrogen use, oral contraceptive use, endocrine disorders and tranquilizer use.On average, women from whom breast fluid was obtained were younger than women from whom breast fluid was attempted but not obtained . In unadjusted and age-adjusted comparisons, being married, a history of pregnancy, younger age at menarche (12 years of age or younger), tranquilizer use, oral contraceptive pill (OCP) use and endocrine problems were associated with an increased ability to obtain breast fluid. Post-menopausal status and exogenous estrogen use were associated with a decreased ability to obtain breast fluid. After age-adjustment, oral contraceptive use was no longer significantly associated with an increased ability to obtain fluid and post-menopausal status was no longer associated with a decreased ability to obtain breast fluid. After multivariate adjustment, age, being married, a history of pregnancy, tranquilizer use and a history of endocrine problems remained positively associated with the ability to obtain breast fluid. In addition, menopausal women who took estrogen were less likely to yield fluid than premenopausal women.Four variables  remained positively associated with the ability to obtain NAF in all analyses. A younger age was consistently associated with a greater ability to obtain NAF in this and other studies. The search for early stage breast cancer, or more importantly, its precursors, is the goal of many researchers hoping to reduce mortality rates from this disease. Current methods of breast cancer early detection include mammography, ultrasound, clinical breast exam (CBE) and breast self-examination (BSE). In addition to these methods, other imaging techniques such as magnetic resonance imaging (MRI) are being explored . AlthougIn order to utilize NAF as both an adjunct to current breast cancer risk assessment tools and/or for breast cancer early detection, it is important to understand factors influencing the obtainment of adequate fluid samples. Past studies have shown four factors consistently associated with an increased ability to obtain breast fluid; age between 35–50 years, earlier age at menarche, non-Asian compared to Asian ethnicity and a history of lactation . Using aAll women were volunteers seen in a private breast clinic by one physician, Dr. Otto Sartorius, who attempted NAF collection on 3413 women. The women came to the clinic for a variety of reasons and NAF was attempted as part of Dr. Sartorius's normal evaluation. The volunteers consisted of women without known breast disease, as well as private clinic patients with known or suspected breast disease . For thiThe nipple aspiration technique utilized on this cohort was developed by Dr. Otto Sartorius . Dr SartStatistical Analysis Software (SAS)  version The women seen at the clinic were between the ages of 15–89 years at the time of breast fluid extraction. Although ethnicity data were only recorded for 16% of the cohort, the geographic area is predominantly Caucasian. Of the ethnicity information known for 16% of this cohort, 89.1% of those were Caucasian (N = 443). Cohort demographics are listed in Table Fluid was obtained from 1314 women (43%) and attempted but not obtained from 1729 women (57%). Overall, the mean age for women who yielded breast fluid was 41.9 years versus 46.5 years for women from whom aspiration was unsuccessfully attempted (p=<.0001). NAF production is lower in women under 30, and then relatively constant for the next 25 years, and then declines in older age women . Fluid was less likely to be obtained from women who used exogenous estrogen Table . After mThis study builds on the work of Wrensch et al.  and otheAlthough being married was positively associated in multivariate analysis with increased ability to obtain fluid in the present study, it was examined in only one other comparison study  and was A previous history of pregnancy was highly significant in the present study in all analyses and this is consistent with findings from Buehring  and PetrIn the present study, postmenopausal women who used estrogen were significantly less likely to yield fluid compared with premenopausal women. Menopausal estrogen use was not found to be significant in any of the previous comparison studies -16 with Tranquilizer use was positively associated with an increased ability to yield breast fluid; however, this factor was not found to be significant in the only comparison study that considered this factor . SartoriThe endocrine disorders studied from the Sartorius cohort produced conflicting findings, given that the disorders included are diabetes and both hyper- and hypo-thyroidism. Wynder et al.  studied Factors which were not found to be significantly associated with breast fluid yield in the present study which are consistent with lack of significance in other studies include menopausal status and oral contraceptive use. Only one study found oral contraceptive use to be significantly related to an increased ability to obtain breast fluid . The folWhile the Sartorius cohort represents a unique group from which to examine important factors related to obtaining NAF, the data do have some limitations. Most notable is that in contrast to other studies, a structured questionnaire was not used to obtain information on potentially relevant factors. Instead, the data were obtained from medical record abstraction. It is possible that there may have been some inconsistency over time in how or whether factors were recorded in the medical records. Also, the cohort included women with suspected breast disease; therefore, the ability to generalize the findings may be limited. All available information regarding the characteristics of the cohort was disclosed in Table Taken together, this and other studies suggest that age is the most consistent factor affecting obtainment of breast fluid. Wrensch et al. ,8 obtainBMI – Body mass indexBSO – Bilateral salpingo-oophorectomyBSE – Breast self-examinationCBE – Clinical breast examMRI – Magnetic resonance imagingNAF – Nipple aspirate fluidOCP – Oral contraceptive pillsPRC – Protocol Review CommitteeSAS – Statistical Analysis SoftwareUCSF – University of California San FranciscoThe author(s) declare that they have no competing interests.KB conceived of the study and set up the design and coordination of the research team. She also reviewed and interpreted all statistical analyses and drafted the original manuscript. MW participated in the study design and helped to draft the manuscript, including reviewing data analysis and providing assistance with revisions. JS performed statistical analyses. All authors have read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


From 1968 to 2002, Singapore experienced an almost three-fold increase in breast cancer incidence. This increase appeared to be different across the three main ethnic groups: Chinese, Malays and Indians. This paper used age-period-cohort (APC) modelling, to determine the effects of age at diagnosis, calendar period, and birth cohort on breast cancer incidence for each ethnic group.This study included all breast cancer cases  in the three ethnic groups, reported to the Singapore Cancer Registry from 1968 to 2002 between the ages 25 to 79. Age-specific fertility rates from the Department of Statistics were used to explore the role of fertility.In the 1970s, Indian women had the highest age-standardized breast cancer but by the mid-1980s the highest rates were seen among the Chinese. Remarkable differences were seen in the age-specific incidence rates by ethnic groups. After age 49, the incidence rates for the Chinese and Malays leveled off whereas it continued to rise in the Indians. While our analyses provided some evidence that an age-drift model described the trend seen in the Indians, age-period-cohort model and age-cohort model had the best fit for the Chinese and Malays aged 25 to 79 respectively. Overall, Chinese and Malay women born in later cohorts were at increased risk of developing breast cancer relative to their counterparts in the earlier cohorts. The three ethnic groups experienced similar changes in their fertility in the 1970s, which likely explained much of the increase in their breast cancer incidence but not the ethnic differences. There was a stronger inverse association between total fertility rate and pre-menopausal breast cancer incidence in the Chinese and Malays than the Indians.The observed dissimilarity among ethnic groups suggests ethnic differences in exposure or response to certain risk factors. It is likely that longer and subtler differences in childbearing trends and other risk factors may further explain these ethnic differences. The rapid rise in female breast cancer incidence in Singapore has been attributed to a birth cohort effect that probably arised from the rapid changes in reproductive and other lifestyle patterns . The curTime trends in breast cancer rates may be attributed to an aging population (age effect), changes occurring at a specific calendar period (period effect) or changes affecting persons born at specific times (cohort effect). Hence an age-period-cohort (APC) model can be used to assess the effects of the three factors on disease rates ,6.Breast cancer incidence rates have been reported to be different between Asian and Caucasian populations ,7,8. TraSingapore is characterised by three main ethnic groups: Chinese, Malays and Indians. These three ethnic groups have remarkable differences in breast cancer incidence in spite of relatively similar changes in reproductive and lifestyle changes. Our primary aim is to explore the trends in the incidence rates of breast cancer in the three ethnic groups through the use of APC modelling and if differential reproductive and lifestyle patterns between the ethnic groups could offer some explanation behind any trends observed.This is a retrospective population-based study performed using data from the Singapore Cancer Registry. The Singapore Cancer Registry was established in 1968 where all medical practitioners and pathology laboratories voluntarily notify the registry of any incident cancers. Staff of the Registry also reviews cancer patient hospital discharges and death certificates against registered cases to ensure completeness of records. The completeness of reporting is high: 96% in the 1970s and close to 100% in the 1990s . In addii) the number of breast cancer cases in 5-year intervals for both age (0 – 80+) and year of diagnosis (1968 – 2002), ii) the 5-year age-specific rates, and iii) the age-standardized rates using world standard population [Data on all breast cancer cases among residents diagnosed between 1968 and 2002, stratified by ethnic group  were obtained from the 5-yearly reports published by the Singapore Cancer Registry . These dpulation . The annpulation .As age at first birth and number of children influence breast cancer incidence, scatterplots of total fertility against cumulative breast cancer incidence rates were used to explore the association of breast cancer with total fertility by birth cohorts. Due to incomplete data, we only present scatterplots for the pre-menopausal women (aged 25 – 49) and five birth cohorts in 2-year interval . The cumulative incidence rates estimate the risk of developing breast cancer for a particular birth cohort of women from age 25 to 49, computed from cancer incidences by single age and year of diagnosis requested from the cancer registry. The total fertility rate (TFR) for a particular birth cohort of women estimates the number of children 1,000 of them would bear. To obtain the TFR by cohort from the annual fertility rates in 5-year age groups, we did a linear interpolation between two midpoints of adjacent age groups to obtain the yearly age-specific fertility rates (i.e. assuming linearity in rates across the two age groups). We also plotted the TFR by period (sum of the age-specific birth rates of women in five-year age groups multiplied by five) that estimates the number of children 1,000 women would bear if the age-specific birth rates of a specified calendar-year reflected their childbearing pattern.APC modelling was used to fit the Singapore breast cancer data. We have breast cancer data in the period 1968 – 2002  but restricted the analysis to women aged 25 – 79  because of too few events in the younger age groups and data quality in the oldest age group. In the modelling procedure, we took a model-building approach [For the full APC model, it is well-known that the linear relationship between age, period and cohort does not allow simultaneous estimation of all three effects unless additional assumptions are imposed ,6. TherePoisson regression with a log-link was used by APC modelling techniques to model the number of breast cancer cases. We adjusted for the total number of person-years for each age-period combination through an offset term . Akaike'The incidence rate ratios (IRR) obtained from APC models were subsequently used to describe the effects associated with calendar period and birth cohort . We firsFrom 1968 to 2002, there were 16,178 breast cancer cases reported to the Singapore Cancer Registry, of which 15,269 were diagnosed in Chinese (85.2%), Malay (9.9%) and Indian (4.9%) women between the ages of 25 to 79. The overall age-standardized incidence rates  ethnic differences in exposure to certain risk factors, or ii) ethnic differences in the response to similar changes in exposure, or iii) both differences in ethnic exposure and responses.These observed ethnic differences in age-specific rates and results from APC modelling suggest that there may be The changes in breast cancer incidence may be related to Singapore's transition from an industrialized country to a developed country, bringing about changes in lifestyle in the population, including late marriages, delay in childbearing and smaller family sizes. This economic transition led to demographic changes such as low population growth rates, accompanied by changes in socioeconomic status such as increase in incomes and women's participation in paid employment; and higher education levels . FurtherFrom recent data on parity, there was a general decline in number of children born per ever-married female. In 1980, Malay women had the highest mean number of children (3.9) compared to the other two ethnic groups (3.4 for Chinese and for Indians). By year 2000, the numbers declined to 3.1 for the Malays, 2.5 for the Chinese and 2.4 for the Indians ,29. The Delayed age at first birth and decreasing parity are possibly important factors governing the changes in breast cancer rates of the Chinese and Malays, and may contribute to the increasing IRRs seen in the birth cohorts after 1930, particularly after 1945. Implementation of the anti-natalist family planning campaign in the early 70s would have had more influence on women in their late teens and twenties, i.e. those who were likely to be born after 1940. While among those who had already given birth we may expect to see lower parity, we would expect to see both delayed first birth and lower parity for the remaining women. This is consistent with the increased rate of change of risk for cohort 1946 – 1950 Figure  and the The impact of delaying first birth and decreasing parity brought by the changes in lifestyle patterns and family planning campaign was probably not as great for the Chinese as it was for the Malays since the Chinese already tended to delay age at first birth and had smaller family sizes relative to the Malays (and Indians) in the 1970s. This is consistent with Figure Although all three ethnic groups experienced delayed age at first birth and decreasing parity, there appears to be ethnic differences in its influence on breast cancer incidence. Decreasing total fertility is found to be associated with increasing cumulative incidence rates in later birth cohorts for Chinese and Malay women aged 25 to 49 .Our preliminary analyses showed ethnic differences in the association between fertility and pre-menopausal breast cancer. As this cohort matures, the ethnic differences in post-menopausal breast cancer can be further examined. However, total fertility rates may not be an adequate index of the changing fertility patterns as it does not capture the changes in age at first birth of a population, which is known to be a strong protective factor for breast cancer. A record-linkage study between birth registry and cancer registry may confirm the ethnic differences associated with fertility.2) has generally been more prevalent among the Malay and Indian women compared to the Chinese women [Besides reproductive factors, height, dietary habits, obesity, duration of breastfeeding, childhood growth, age at menarche and menopause are some other possible environmental risk factors whose distribution may vary across the ethnic groups, and could also influence breast cancer risk ,11,31. Ose women -34. Withse women . Howeverse women . Thus wise women ,37. As mse women .This ecological study showed that the temporal trends in breast cancer incidence are different among the Chinese, Malays and Indians in Singapore in spite of fairly similar dramatic declines in fertility rates. Longitudinal cohort data with detailed information on reproductive patterns and other predictive factors will help our understanding of the ethnic differences in the trend of breast cancer incidence in Singapore.The author(s) declare that they have no competing interests.XS, R.AA and KSC conceived of the study, and participated in its design and coordination. XS, R.AA, CST and MR carried out the statistical analysis. SW, DLMG, PH and KSC contributed to the epidemiological aspects. All authors contributed to the writing of the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


This article aims to explain the reasons for the remarkable change in age of breast cancer occurrence in the Swiss canton of Geneva.We used population-based data from the Geneva cancer registry, which collects information on method of detection, stage and tumour characteristics since 1975. For patients diagnosed between 1997–2003, we obtained additional information on use of hormone replacement therapy from a large prospective study on breast cancer. Using generalized log linear regression analysis, we compared age-specific incidence rates with respect to period, stage, oestrogen receptor status, method of detection and use of hormone replacement therapy.In the periods 1975–1979 and 1985–1989, breast cancer risk increased with age, showing the highest incidence rates among women aged ≥ 85 years. From 1997, the age-specific incidence curve changed completely (p < 0.0001), showing an incidence peak at 60–64 years and a reduced incidence among elderly women. This incidence peak concerned mainly early stage and oestrogen positive cancers and was exclusively observed among women who ever used hormone replacement therapy, regardless whether the tumour was screen-detected or not.The increasing prevalence of hormone replacement therapy use during the 1990s could explain the important change in age-specific breast cancer incidence, not only by increasing breast cancer risk, but also by revealing breast cancer at an earlier age. In Western countries, breast cancer is the most frequent cancer among women and its incidence is still increasing. Several factors have contributed to this increase. Screening mammography temporarily increases incidence by discovering incident, as well as prevalent cases. Higher exposure to breast cancer risk factors, such as low and late parity, alcohol intake and use of hormone replacement therapy (HRT) ,2 could The Swiss canton of Geneva has one of the highest incidence rates of breast cancer in the world . OpportuUse of HRT was very common in Geneva where over 50% of women aged 45–59 years used HRT in 1996 .In this study, we describe changes in age-specific breast cancer incidence in Geneva and examine possible reasons for these changes.We used data from the Geneva cancer registry, which records information on all incident cancer cases that occur in the canton . The registration is based on several sources of information and is very accurate, as attested by its low percentage (<2%) of cases recorded from death certificates only . All hosFor every cancer case, we record sociodemographic characteristics, method of detection of the tumour, tumour characteristics, stage of disease at diagnosis, treatment, survival status, and cause of death.The study included all 8045 invasive breast cancer cases diagnosed between 1975–2003 among the resident population. Method of detection was categorized as mammography screening vs. other. Stage at diagnosis was based on the Tumour, Node and Metastasis TNM classification system . Stages Additional information on the patient's use of hormone replacement therapy (coded as ever vs. never user) was provided by a large study on risk factors for breast cancer . This stThis study included 400 breast cancer patients aged 50–74 years, diagnosed between 1997–2000 in the canton of Geneva and they were individually linked to the cancer registry data set. These women represented 46% of all breast cancer patients in this age group recorded at the cancer registry during the same period.We calculated age-specific breast cancer incidence rates per 100'000 by five-year age groups for four different periods . Differences between age-specific incidence curves were analyzed by generalized log linear regression analysis . We compFigure Figure This study shows a remarkable change in age-specific breast cancer incidence in Geneva. In developed countries, other than Japan, the typical age incidence curve of breast cancer described a progressive increase of risk with age, with a slope down around the menopause age, called the Clemmensen's hook. This typical curve by age is no longer observed in Geneva, where the risk of developing breast cancer does not increase with advancing age anymore. The highest breast cancer risk is now observed among women aged 60–64 years with strongly decreased risks among older women. The incidence peak of women aged 60–64 years was only observed in early stage disease and in oestrogen receptor positive tumours. In the subgroup of women for whom information on HRT use and mammography screening was available, the incidence peak was only present among women who ever used HRT.The prevalence of screening and HRT use are high in our study population and we had the opportunity to examine the respective effects of each of these factors on the change in age of breast cancer occurrence. However, we realise that our study has several shortcomings. Information on HRT use was available only for approximately 50% of the women diagnosed in the period 1997–2000. Although these women were recruited in the context of a population-based case-control study, they might not be completely representative of all breast cancer patients. We compared them with the other 50–74 years old breast cancer patients recorded at the cancer registry and they appeared to be similar in terms of age, social class, stage distribution and tumour characteristics. Therefore, we believe that selection bias does not explain the fact that the new age distribution is only present among women who used HRT, irrespective of their mammography screening status. In addition, we have no information on the duration of HRT utilisation, on HRT cessation, or on time elapsed since the stop of HRT. It is therefore not possible to evaluate if the incidence peak occurred among current or ex-users of HRT. Also, we were not able to evaluate incidence patterns according to type of HRT.To our knowledge, only three studies have reported important changes in age-specific breast cancer incidence -16. In MIn our population, we believe that both HRT and, to a lesser extent, mammography screening explain the change in incidence pattern. Mammography screening advances diagnosis by detecting breast cancer at an early pre-clinical stage. This can explain why the change in pattern concerned mainly the early stages. In this study however, we did not observe the typical incidence peak among never users of HRT with screen-detected tumours. In fact, only ever users of HRT showed an incidence peak at 60–64 years, regardless whether the tumour was screen-detected or not.It has been suggested that HRT particularly increases the risk of oestrogen receptor positive tumours . In our Recent studies showed that current users of HRT are at increased risk of developing breast cancer . These studies generally involved patients between 50–65 years and the effect of ever use of HRT on breast cancer incidence among elderly women (aged ≥ 75 years) has, to our knowledge, never been examined.Our results suggest that HRT use might not only increase breast cancer risk among middle-aged women. It may also bring forward the clinical appearance of breast cancer by several years by stimulating the growth of oestrogen sensible lesions and by inhibiting spontaneous slow down of tumours when sex hormone levels drop during menopause. This hypothesis is supported by a previous study, which showed that postmenopausal women using HRT were on average five years younger upon breast cancer diagnosis than never users [20].Use of HRT could have eliminated the temporary protective effect of menopausal oestrogen depletion.The results of this study suggest that HRT may not only increase the breast cancer risk among middle-aged women, but that it may also bring forward the clinical appearance of breast cancer. To get more insight in the complex relationship between age-specific breast cancer incidence, mammography screening and HRT, it could be useful to perform international comparisons of cancer registry data covering populations with different mammography screening policies and different prevalences of HRT use. The change in incidence pattern could have an impact on breast cancer mortality. If the increase is "only" anticipation in age at onset, we can expect that the subsequent mortality will remain stable. However, there is also an anticipation of stage and that could lead to a future lowered mortality. If it is a true increase in risk, then we can expect a subsequent possible increase in mortality.The author(s) declare that they have no competing interests.Christine Bouchardy, Alfredo Morabia, Helena M. Verkooijen, Gérald Fioretta and Peter Schäfer designed the study. Alfredo Morabia, Yves Wespi, Gérald Fioretta and Peter Schäfer were involved in data collection. Christine Bouchardy, Gérald Fioretta and Helena M. Verkooijen performed the statistical data analyses. All authors were involved in the interpretation of the data. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


This report presents the first meta-analysis of results from epidemiological studies that have investigated parity, age at first birth, breastfeeding, and age at menarche in relation to ERWe calculated summary relative risks (RRs) and corresponding 95% confidence intervals (CIs) using a fixed effects model.+PR+ cancer by 11% , and women who were in the highest age at first birth category had, on average, 27% higher risk of ER+PR+ cancer compared with women who were in the youngest age at first birth category . Neither parity nor age at first birth was associated with the risk of ER-PR- cancer . Breastfeeding and late age at menarche decreased the risk of both receptor subtypes of breast cancer. The protective effect of late age at menarche was statistically significantly greater for ER+PR+ than ER-PR- cancer .Each birth reduced the risk of EROur findings suggest that breastfeeding (and age at menarche) may act through different hormonal mechanisms than do parity and age at first birth. A review published in 2004 " and "(ER or PR).mp ". We identified additional studies by tracking the references in all identified articles. We noticed that the studies published before 1995 all defined their receptor subtypes according to either ER or PR status and that most of them had a hospital-based study design, whereas most of studies published since 1995 used joint ER/PR status to define receptor subtypes and had a population-based study design. Using joint ER/PR status could reduce the chance of including any tumors in which one of the receptor statuses was mislabeled. Therefore, for inclusion into this meta-analysis, the identified articles have to have estimates of relative risk (RR) for ERo cohort , five poo cohort ,11-14, ao cohort ,16 that o cohort   and their 95% confidence intervals (CIs) for highest versus lowest category of parity, age at first birth, breastfeeding, and age at menarche. For this analysis, we used RR to refer to any RR measure. Two studies presented their results by subgroups only, either by menopausal status  or by agFor parity, we also extracted or computed study-specific trend estimates. One study provided the estimates in the original publication ; we compTo summarise the results by menopausal status, we did as follows. We accepted the definitions used in three studies ,11,13. Tp ≥ 0.10) [+PR+ and ER-PR- subtypes, we used the inverse of the average variances of the two subtypes as the weighting variable. This is necessary for the calculated summary values to be directly comparable. In Tables We used fixed effects models to calculate summary RRs for all studies combined, by type of study design and menopausal status , because we did not detect statistically significant heterogeneity of effects between studies ( ≥ 0.10) . To ensu2 (with 1 degree of freedom), this value is squared and then divided by its variance, which is the inverse of the sum of the inverses of the pooled variances across studies. Potential heterogeneity in effect by study design and by menopausal status was examined by using standard homogeneity tests. All p values reported for homogeneity are two-sided, and p values less than 0.05 were considered statistically significant.To test for potential heterogeneity in risk by receptor status, we first calculated the average variance weighted difference in the log RRs for the two receptor status groups. The variance for each difference in log RRs is estimated by the sum of their variances. The differences are divided by their pooled variance, these weighted values are summed across studies, and the total is divided by the sum of the inverses of these pooled variances. To create a test statistic, χp value of less than 0.10.Using Egger's regression asymmetry test, we assessed the possibility of publication bias . This anAll analyses were performed using the Stata statistical software .+PR+ and 80 to 1,081 for ER-PR-. On average, the number of ER-PR- breast cancer cases involved in the analyses was approximately 46% of that of ER+PR+ cancer cases. All studies considered confounding in their analyses although the confounders included in the models varied by study , and the p value for homogeneity between ER+PR+ versus ER-PR- cancer was less than 0.001 . The difference in effects between ER+PR+ and ER-PR- was statistically significant (p for homogeneity between ER+PR+ and ER-PR- cancer = 0.010). The summary RR for ER+PR+ cancer appeared greater among postmenopausal than premenopausal or young women , but the difference was not statistically significant (p = 0.211).Nine studies were included in the meta-analysis of age at first birth and breast cancer risk by ER/PR status Table . Women i+PR+ and ER-PR- cancer  (Table p = 0.071 for ER+PR+ and p = 0.054 for ER-PR- cancer). One reason for this discrepancy could be that the two hospital-based studies included a small number of cases and therefore had insufficient power to find any effect.The summary RRs from the seven studies of breastfeeding show that breastfeeding was associated with reduced RRs of both ER 0.91 0.8–1.00 for+PR+ and ER-PR- cancer . The protective effect of late age at menarche was statistically significantly greater for ER+PR+ than ER-PR- cancer (p for homogeneity between ER+PR+ and ER-PR- cancer = 0.006).Nine studies were included in the meta-analysis of age at menarche and breast cancer risk by ER/PR status Table . The ovep > 0.10).We found no evidence of publication bias in results across all studies for the factors we reviewed . We therefore think it is unlikely that the difference in association by receptor status is due to residual confounding by age.We also considered whether the different association by receptor status could be caused by residual confounding by age given that the ratio of ERwith age ,21. We twith age . We founThe main source of hormone receptor information for studies that we reviewed was medical records. Although we assume that the majority of laboratories have used immunoassays since 1995, we could not exclude the possibility that the assays and cutoffs for determining ER and PR status differed across studies. However, we believe that any such inconsistencies would be unlikely to cause the observed associations and, if anything, that they would bias the RR estimates toward the null value.-PR- cancer [Some data suggest that compared with Caucasian women, African-American women are more likely to develop ER- cancer . We were- cancer . HoweverThe differences between the comparison and reference categories for age at first birth varied substantially across the studies, ranging from 1 ,12,14,16+PR+ but not ER-PR- breast cancer suggest that their effects influence risk predominantly through hormonal mechanisms that involve estrogen and progesterone. The effects of these hormones on breast tissue depend upon the amount of both hormones and their specific receptors [The protective effects of a greater number of births and an early age at first birth against EReceptors -25. A greceptors -28, (b) eceptors , or (c) eceptors .+PR+ and ER-PR- subtypes, although menarche had greater protective effects against ER+PR+ than did ER-PR- cancer. This seems to be inconsistent with the hypothesis that these factors act through estrogen and progesterone mediated by their respective receptors [+ progenitor cells are exposed to estrogen, they produce paracrine signals that cause neighbouring populations of ER- cells to proliferate [Contrary to expectations, breastfeeding and late age at menarche protected against both EReceptors -25. Howeliferate . Thus, o+PR+ breast cancer whereas breastfeeding and late age at menarche protect against both ER+PR+ and ER-PR- breast cancer. Our findings suggest that breastfeeding (and age at menarche) may act through different hormonal mechanisms than do parity and age at first birth.Our quantitative overview shows that parity and early age at first birth protect only against ERCI = confidence interval; ER = estrogen receptor; PR = progesterone receptor; RR = relative risk.The authors declare that they have no competing interests.HM conducted the literature search and assembled the data, conducted the data analysis, and drafted the manuscript. LB revised the manuscript. MCP supervised the data analysis and revised the manuscript. GU provided advice on the literature search, data collection, and analysis and revised the manuscript. All authors have read and approved the final draft.


Modeling the relationship between age and mortality for breast cancer patients may have important prognostic and therapeutic implications.Data from 9 registries of the Surveillance, Epidemiology, and End Results Program (SEER) of the United States were used. This study employed proportional hazards to model mortality in women with T1-2 breast cancers. The residuals of the model were used to examine the effect of age on mortality. This procedure was applied to node-negative (N0) and node-positive (N+) patients. All causes mortality and breast cancer specific mortality were evaluated.The relationship between age and mortality is biphasic. For both N0 and N+ patients among the T1-2 group, the analysis suggested two age components. One component is linear and corresponds to a natural increase of mortality with each year of age. The other component is quasi-quadratic and is centered around age 50. This component contributes to an increased risk of mortality as age increases beyond 50. It suggests a hormonally related process: the farther from menopause in either direction, the more prognosis is adversely influenced by the quasi-quadratic component. There is a complex relationship between hormone receptor status and other prognostic factors, like age.The present analysis confirms the findings of many epidemiological and clinical trials that the relationship between age and mortality is biphasic. Compared with older patients, young women experience an abnormally high risk of death. Among elderly patients, the risk of death from breast cancer does not decrease with increasing age. These facts are important in the discussion of options for adjuvant treatment with breast cancer patients. In many clinical situations, age is an important determinant of treatment decision in breast cancer. For example, after mastectomy, patients with T2 tumors and one to three positive nodes are at high risk of isolated loco-regional recurrences. Authors have advocated the routine use of postmastectomy radiotherapy in those patients who have T2 tumors and who are younger than 45 years . In anotAt the other end of the age spectrum, breast cancers in elderly patients have been considered by some authors to exhibit a less aggressive behavior than in younger patients ,5. OtherThere are still controversial issues about the relationship between age and prognosis in breast cancer. Detailed analysis would be useful in order to provide more insight into this relationship. In the present study, we used proportional hazards to model the survival of T1-2, node-negative (N0) and node-positive (N+) breast cancer patients. Outcomes which we considered included all-cause mortality and cancer specific mortality from breast cancer. The primary aim of the study is to present how age relates with the risk of death. The secondary objective is to search for a simple algebraic representation of this relationship.The Surveillance, Epidemiology, and End Results Program (SEER) of the United States collected data about the incidence of cancer and related matters from 11 population-based registries . The datIn order to verify the linearity of the continuous variables, the martingale residuals (differences between observed and expected numbers of events) were used. The martingale residuals were examined by a non-parametric smoothing  against the quantitative covariates of interest. The smoothing used a Poisson regression implementation of generalized additive model (GAM) . The GAMThe analysis was applied first to node-negative cases ("training set") in order to find a simple expression of the functional form which relates age to mortality. The functional form obtained from node-negative cases was then applied to node-positive cases . In addition to the validation with the same transformation which was obtained for node-negative patients, a further iterative search was performed in order to improve the fit for node-positive patients.This analysis was applied also to a European dataset, the German Breast Cancer Study Group (GBSG-2), in which the outcome studied was disease-free survival . From a The statistical analyses were performed with Splus  statistical software. Parametric fitting of curves used TableCurve 2D .There are 83,804 T1-2 cases  available for analysis from the SEER database. Table all causes, for N0 and N+ patients, respectively. Both curves have similar U-shapes. The mortality is lowest for patients about 50 years of age at diagnosis. The mortality increases the farther away from 50 years of age at diagnosis, for both younger and older patients.Figures (age) and a quasi-quadratic term |age-50|1.5, i.e. age+ |age-50|1.5, provides a good fit and passes the test of linearity  in N0 patients . The further age at diagnosis is from the age at menopause, the more prognosis is influenced by the quasi-quadratic component.The age transforms suggest two components in the effect of age. One component is linear  and corresponds to a natural increase in mortality with each year of age. The other component is quasi-quadratic and is centered around age 50. It contributes to an increased risk of mortality as age increases beyond 50. It suggests a hormonally related process, not pre- versus post-menopausal, but The results display a complex functional form of the effect of age on mortality. The curves clearly highlight the biological anomaly that younger patients experience the same relative mortality risk from all causes as do older patients. Figures The marked increase in mortality risk at older ages is attributable to the increased risk of death from causes other than breast cancer (co-morbidity). It should be noted that breast cancer does not become less virulent in older patients. An increase in the risk of death from breast cancer associated with older age was observed both in N0 and in N+ patients Figures  and 4.2 = 31.744, 3 degrees of freedom, P < 0.000001). The age transforms improved the linearity for the age variable, and also improved the proportional hazards model , Hollaender found that all methods showed a decrease in risk with increasing age up to 45–50 years . A slighRegarding the proportional hazards assumption, Hollaender noted that assuming a linear risk function, a small correlation value rho of 0.147 was obtained . Our resOur results are also in keeping with a closely related investigation of the SEER data in which a group of 4,616 patients 35 years old or younger was compared to a group of 20,319 patients aged 50–55 years . The autIn order to try to understand the biphasic mortality, we looked at hormonal status and treatments of the patients. The age of 50 corresponds to the menopause. A large proportion of younger women were estrogen receptor (ER) negative  declare that they have no competing interests.PT, VVH: writing the manuscript; SJL, VVH: data analysis;GCs, JVDS, GV, MR, MV, GS: concept, design, and drafting the manuscript.All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:Check of the Cox proportional hazards assumption.Click here for fileOutput of Cox, GAM and cox zph.Click here for fileGBSG-2 data set.Click here for file


TYMS 5'-untranslated enhanced region (TSER) and a 6-bp deletion/insertion in the TYMS 3'-untranslated region (TS 3'-UTR)] were suggested to be correlated with alteration of thymidylate synthase expression and associated with cancer risk.Accumulative evidence suggests that low folate intake is associated with increased risk of breast cancer. Polymorphisms in genes involved in folate metabolism may influence DNA methylation, nucleotide synthesis, and thus individual susceptibility to cancer. Thymidylate synthase (TYMS) is a key enzyme that participates in folate metabolism and catalyzes the conversion of dUMP to dTMP in the process of DNA synthesis. Two potentially functional polymorphisms [a 28-bp tandem repeat in the TYMS gene are associated with risk of breast cancer, we genotyped these two polymorphisms in a case-control study of 432 incident cases with invasive breast cancer and 473 cancer-free controls in a Chinese population.To test the hypothesis that polymorphisms of the TS3'-UTR (1494del6) genotype frequencies were significantly different between the cases and controls (P = 0.026). Compared with the TS3'-UTR del6/del6 wild-type genotype, a significantly reduced risk was associated with the ins6/ins6 homozygous variant genotype  but not the del6/ins6 genotype . Furthermore, breast cancer risks associated with the TS3'-UTR del6/del6 genotype were more evident in older women, postmenopausal subjects, individuals with a younger age at first-live birth and individuals with an older age at menarche. However, there was no evidence for an association between the TSER polymorphism and breast cancer risks.We found that the distribution of TS3'-UTR del6 polymorphism may play a role in the etiology of breast cancer. Further larger population-based studies as well as functional evaluation of the variants are warranted to confirm our findings.These findings suggest that the  A wide d breaks -10 and ad breaks ,12 that Many enzymes are involved in folate metabolism, among which, thymidylate synthase (TYMS) catalyzes the conversion of deoxyuridine monophosphate (dUMP) to deoxythymidine monophosphate (dTMP), using the 5, 10-methylenetetrahydrofolate as a methyl donor . This coTYMS gene is located in chromosome 18p11.32. A tandem repeat polymorphism has been identified in the 5'-UTR enhancer region of the TYMS promoter (TSER), the immediate upstream of the ATG codon initiation start site, which contains triple (TSER 3R) or double (TSER 2R) repeats of a 28-bp sequence as well as several rare alleles containing 4, 5 or 9 repeats [In vitro and in vivo studies showed that TYMS expression was TSER genotype-dependent and that the 3R allele was associated with a higher TYMS expression level [TYMS gene (TS3'-UTR del6), has also been identified, which may have an effect on the TYMS mRNA stability and translation [TYMS variants are thought to be functionally relevant and are hypothesized to be associated with risk of breast cancer. To test this hypothesis, we performed genotyping analyses for TS 5'-UTR TSER and TS 3'-UTR del6 polymorphisms in a case-control study of 432 incident cases with invasive breast cancer and 473 cancer-free controls in a Chinese population.The  repeats . In vitron level ,17. Recein situ breast cancer (DCIS), metastasized cancer originated from other organs, and previous radiotherapy or chemotherapy were excluded. Cancer-free controls were randomly selected from a pool of 10,500 individuals who participated in a community-based screening program for non-infectious diseases conducted in Jiangsu Province during the same time period as the cases were recruited. These control subjects had no self-reported cancer history and were frequency-matched to the cases on age (± 5 years) and residential areas. Each subject was scheduled for an interview after an informed consent was obtained, and a structured questionnaire was administered by interviewers to collect information on demographic data, menstrual and reproductive history, family history of cancer (any kind of cancers in first-degree relatives) and environmental exposure history. After the interview, an approximately 5-ml venous blood sample was collected from each subject. The study was approved by the institutional review board of Nanjing Medical University.This hospital-based case-control study consisted of 432 patients who received surgical treatment for solitary mammary lumps, which were histopathologically diagnosed as invasive breast cancer, and 473 cancer-free controls. All subjects were genetically-unrelated ethnic Han Chinese and were from Nanjing city and surrounding regions in Jiangsu Province. The breast cancer patients were consecutively recruited between January 2004 and July 2005 at the Cancer Hospital of Jiangsu Province, the First Affiliated Hospital of Nanjing Medical University and the Nanjing Gulou Hospital, Nanjing, China. Those patients who had any previous cancers, TYMS [TSER locus and 5'-CAAATCTGAGGGAGCTGAGT-3' (forward) and 5'-CAGATAAGTGGCAGTACAGA-3' (reverse) to genotype the TS 3'-UTR del6 locus. For the TSER polymorphism, 243 bp (i.e. 3R), 215 bp (i.e. 2R), 271 bp (i.e. 4R) and 299 bp (i.e. 5R) fragments were identified and separated on 3% agarose gels. For the TS 3'-UTR del6 polymorphism, 152 bp (i.e. del6) or 158 bp fragments (i.e. ins6) were amplified and then digested by Dra I . The variant (ins6) allele produces two fragments of 88 and 70 bp, while the wild-type (del6) allele produces a single 152-bp fragment. Genotyping was performed without knowing the subjects' case or control status, and approximately equal number of cases and controls were assayed in each 96-well PCR plate with a positive control of a DNA sample with a known heterozygous genotype. If a consensus on the tested genotype was not reached, two research assistants independently performed the repeated assays to achieve 100% concordance. More than 10% of the samples were repeated using the same assay and the results were 100% concordant.By centrifugation of 5-ml whole blood, genomic DNA was extracted from the leukocyte pellet obtained from the buffy coat of each blood sample. We used the previously-described genotyping assays for the two loci of TYMS . BrieflyTSER and TS 3'-UTR del6 between the cases and controls were evaluated using the χ2 test  and student t-test (for continuous variables). The associations between TSER and TS 3'-UTR del6 variant genotypes and breast cancer risk were estimated by computing ORs and their 95% CIs from both univariate and multivariate logistic regression analyses. Linkage disequilibrium (LD) between the two loci of TYMS was estimated using the EH algorithm available online [TYMS genotypes. Hardy-Weinberg equilibrium was tested by a goodness-of-fit χ2 test to compare the observed genotype frequencies with the expected frequencies among the control subjects. All the statistical analyses were performed with Statistical Analysis System software .Differences in demographic characteristics, selected variables, and frequencies of the genotypes, alleles, and haplotypes of e online . We usede online  to inferTSER and TS 3'-UTR del6 alleles/haplotypes of the 432 breast cancer cases and 473 cancer-free controls are summarized in Table p value with the student t test. Compared with control subjects, breast cancer patients had a lower age at menarche (P <0.0001) and a higher age at having first live birth (P <0.0001). Among the 498 postmenopausal subjects, the menopausal age of breast cancer cases was lower than that of the controls (P = 0.02). There were 113 (26.2%) breast cancer cases and 95 (20.1%) controls reported a family history of cancer, which was associated with a significantly increased risk for breast cancer .The selected characteristics and distributions of TSER 2R and TS 3'-UTR ins6 were 0.204 and 0.331, respectively, in the controls, while they were 0.202 and 0.310, respectively, in the cases, and the differences were not statistically significant (P = 0.987 for TSER and P = 0.347 for TS 3'-UTR del6) . There were four haplotypes derived from the observed TYMS genotypes, but there was no significant difference in the distribution of haplotypes between the cases and controls (P = 0.863) . The TSER genotype frequencies were 64.5% (3R/3R), 30.2% (2R/3R) and 5.3% (2R/2R), respectively, among controls, which were not significantly different from those of the cases  (P = 0.999). Multivariate logistic regression analyses revealed that both 2R/3R and 2R/2R genotypes were not significantly associated with breast cancer risk , compared with the TYMS 3R/3R genotype, (Table TS 3'-UTR del6 genotype were 45.7% (del6/del6), 42.5% (del6/ins6), and 11.8% (ins6/ins6), respectively, in control subjects and 44.9% (del6/del6), 48.1% (del6/ins6), and 6.9% (ins6/ins6), respectively, in breast cancer patients, and the overall difference was statistically significant (P = 0.026). Compared with TS 3'-UTR del6/del6 wild-type genotype, a significantly reduced risk was associated with ins6/ins6 homozygous variant genotype , but it was not related with del6/ins6 heterozygous genotype . , postmenopausal women , individuals with a younger first live birth age  and individuals with a later menarche age   that we previously reported [et al [TSER variant that homozygous triple repeat was twice as common in Chinese subjects (67%) as in Caucasians (38%). Similarly, we found that the allele frequency of TS 3'-UTR ins6 was 0.331 among the 473 control subjects, which is also consistent with what was reported in Chinese populations , but itth China ,30,34.TYMS gene and similar allele frequencies of our controls to those reported from other Asian populations suggests that the selection bias would not be substantial in terms of genotype distribution. Finally, although we have 94% statistical power to achieve the main effect of the study, we have very limited power in the stratified analyses because of the small sample size of the subgroups. Therefore, large well-designed studies are warranted to further confirm our findings.The primary shortcoming of this study is the lack of data on detailed dietary intake of folate, plasma or erythrocyte folate levels and its precursors or metabolites such as homocysteine, which limited our inquiry into potential gene–nutrient interactions in breast carcinogenesis. Furthermore, like all other hospital-based case-control studies, inherited biases may lead to spurious findings or false positive results. Because our cases were from hospitals and the controls were from the community, the study subjects may not be fully representative of the general population. However, we applied rigorous epidemiological designs by limiting factors of selecting subjects and performed further statistical adjustment to minimize these potential biases. The agreement with Hardy-Weinberg equilibrium of these two loci of TS 3'-UTR ins6/ins6 genotype may contribute to the etiology of breast cancer. Studies in diverse ethnic populations and simultaneous measurements of detailed data of the TYMS genotypes as well as its phenotypes, accompanied by multiple SNPs of other important one-carbon metabolism genes and information of folate intake are warranted to further elucidate gene-gene and gene-environment interactions in susceptibility of breast cancer.Our study provides some new evidence that the TSER, thymidylate synthase in the 5'-untranslated enhanced region; TS3'-UTR, thymidylate synthase in the 3'-untranslated region; OR, odds ratio; CI, confidence interval; SNP, single nucleotide polymorphism; PCR – RFLP, polymerase chain reaction – restriction fragment length polymorphism.TYMS, thymidylate synthase; The author(s) declare that they have no competing interests.H. S., X. W., F. C. and X. Z. were responsible for the study design, data analyses and interpretation. J. T., J. Q., S. W., X. W. were principally responsible for patients recruitment with assistance from J. G., G. J., J. L. and W. C.. X. Z. and J. G. were responsible for the laboratory work with assistance from G. J. and W. C.. The report was prepared by X. Z. and Z. H., proof read by Q. W. and H. S.. H. S. made the decision to submit the paper for publication and supervised the whole process of the study. All the authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here: