Smokeless tobacco is often referred to as a major contributor to oral cancer. In some regions, especially Southeast Asia, the risk is difficult to quantify due to the variety of products, compositions (including non-tobacco ingredients) and usage practices involved. In Western populations, the evidence of an increased risk in smokeless tobacco users seems unclear, previous reviews having reached somewhat differing conclusions. We report a detailed quantitative review of the evidence in American and European smokeless tobacco users, and compare our findings with previous reviews and meta-analyses.Following literature review a meta-analysis was conducted of 32 epidemiological studies published between 1920 and 2005 including tests for homogeneity and publication bias.Based on 38 heterogeneous study-specific estimates of the odds ratio or relative risk for smokeless tobacco use, the random-effects estimate was 1.87 . The increase was mainly evident in studies conducted before 1980. No increase was seen in studies in Scandinavia. Restricting attention to the seven estimates adjusted for smoking and alcohol eliminated both heterogeneity and excess risk . Estimates also varied by sex  and by study design  but more clearly in studies where estimates were unadjusted, even for age. The pattern of estimates suggests some publication bias. Based on limited data specific to never smokers, the random-effects estimate was 1.94 (0.88–4.28), the eight individual estimates being heterogeneous and based on few exposed cases.Smokeless tobacco, as used in America or Europe, carries at most a minor increased risk of oral cancer. However, elevated risks in specific populations or from specific products cannot definitely be excluded. Elsewhere, products used are likely to differ in toxicity, and their associated risk, especially in Southeast Asia, is difficult to quantify due to the variety of products, compositions (including non-tobacco ingredients) and usage practices involved. Our objective was to conduct a detailed analysis of published evidence on the association of American and European smokeless tobacco and oral cancer, and to calculate a pooled effect estimate. Compared with the reviews of Rodu and Cole  and CritRW identified reports from a systematic search, not limited by period or language, of MEDLINE, EMBASE, CANCERLIT and TOXLINE through March 2005. The main searches used combinations of the terms "smokeless tobacco", "chewing tobacco", "snuff", and "snus" for exposure, and "oral cancer" and "mouth neoplasms" for outcome. Specific oral cancer sites corresponding to ICD-10 codes C00 to C14 [Inclusion criteria: published in peer reviewed journal or publicly available, based on research on humans, of cohort or case-control design, study location specified, any form of oral cancer as the outcome, and chewing tobacco, orally used moist snuff or unspecified smokeless tobacco as the exposure. They also had to satisfy the Exclusion criteria: sample included in a more complete or recent study, conducted in an Asian population, no control group, inappropriate design , insufficient power (less than five expected exposed cases), appropriate risk estimates and CIs not reported and cannot be computed from the available data.All reports had to satisfy the following Having completed this process, RW then compared the selected studies with those of a previous unpublished review conducted in 2002 by PNL based on MEDLINE, reference lists and reviews.From each report details were abstracted relating to the study , the exposure , the outcome , and issues relating to analysis .To avoid bias, it would be advantageous to restrict attention to estimates of effect size (OR or RR) and of precision (95% CI) based on subjects who have never smoked. However, such results are rarely available, and are often based on very limited numbers of cases. Our main analyses are therefore based on estimates derived from the whole population, smokers and nonsmokers combined, except for studies where the only available results are specifically for smokers or for nonsmokers. However, estimates of effect size and precision specifically for never smokers were also extracted for an additional analysis (see the end of the results section).Where possible, separate estimates were obtained for male and female subjects and for different types of smokeless tobacco; chewing, snuff or smokeless (unspecified or chewing/snuff). Estimates for ever exposure were preferred to estimates for current exposure. Adjusted estimates were preferred to crude estimates and where multiple adjusted estimates were available, that adjusted for the most potential confounding variables was used.In many studies, the required estimates of effect size and precision were not given by the authors and had to be estimated. Crude estimates were derived from the relevant 2 × 2 table using standard methods, with, where necessary, numbers estimated from proportions given numerically or graphically and/or combined over types of oral cancer or levels of exposure. Independent crude estimates were combined over strata by fixed-effect meta-analysis  to obtai2 statistic [For effect size estimate, the standard error of its logarithm was calculated from its reported or estimated CI, assuming that the effect size was log-normally distributed. The logarithms of the effect sizes and their corresponding standard errors formed the data points for fixed-effect and random-effects meta-analyses . Separattatistic .For selected meta-analyses, a forest plot is shown. Each estimate is shown as a rectangle, with its area proportional to its weight. The CI is indicated by a horizontal line. The data are plotted on a logarithmic scale so that the estimate is centred in the CI. Also shown in the plot are the actual values of the estimate and its CI and weight. Results from a random effects meta-analysis are also shown. The combined estimate is presented as a diamond with the width corresponding to the CI, and the estimate as the centre of the diamond.a priori by type of smokeless tobacco, sex, study design, study location, study period, and by whether or not the estimate was reported by the author or was derived by us. Between-subgroup heterogeneity was assessed by a chi-squared test [For the "all" types set of estimates, sensitivity analyses were carried out by eliminating those estimates that were only for smokers or only for nonsmokers, and also by replacing those estimates that were calculated approximately assuming that there were no joint users of chewing tobacco and snuff by estimates for chewing tobacco in US studies and for snuff in Swedish studies. Heterogeneity was also investigated further, separately for estimates unadjusted for any factor and estimates adjusted for smoking. Subgroups were defined red test . To descAdditional meta-analyses were carried out for the set of estimates for smokeless tobacco use among never smokers.Publication bias was assessed by a funnel plot, in which the logarithm of the effect size was plotted against its precision ,26.1313 potentially relevant studies were identified by the literature search, but only 63 appeared possibly of value from initial assessment. These underwent more detailed examination and 32 met the inclusion and exclusion criteria. Of the 31 studies that did not, ten were excluded due to insufficient quantitative information ,27-35, fTable Not all studies were primarily concerned with the relationship between smokeless tobacco and oral cancer. Consequently, study-specific inclusion and exclusion criteria varied, as did criteria for matching of controls. Of the 29 case-control studies, seven  selected controls without smoking-related diseases, five  selected controls without any cancer or without oral and other specific cancers, nine  selected patient controls without oral cancer, and eight  selected population controls essentially without restriction.Table Table Overall, Table As described in Table Overall 16,540 cases of oral cancer were used in the present analysis. Of these, 222 were in three cohort studies , 5,580 were in 19 case-control studies with hospital controls , 10,704 were in nine case-control studies with population controls , and 34 were in the one case-control study with both types of control (study 7). The first, second and third quartiles of the number of cases considered were 94, 200, and 387. The smallest study had 14 cases (study 19) and the largest 6,976 . Excluding study 21, where numbers of controls were not available, a total of 29,009 controls were considered in the 28 case-control studies, of which 17,313 were hospital, 11,536 population and 160 mixed.Diagnoses of oral cancer were reported as histologically confirmed in 19 studies  and as confirmed for some of the cases by four studies . Histological confirmation was not reported by six studies , and death certificates  were used in three studies .Figure Table Based on the 38 estimates for all types of smokeless tobacco shown in Figure When only those estimates adjusted for smoking are considered, the heterogeneity is markedly reduced, and the fixed-effect estimate also reduces, to 1.31. Since significant heterogeneity remains, the random-effects estimate of 1.35 is the preferred overall estimate. When the analysis is further restricted to estimates adjusted for both smoking and alcohol consumption, no heterogeneity is seen. Moreover, the point estimate is close to unity . The lack of detectable risk when these factors are accounted for seems consistent with an important confounding effect of smoking and alcohol, both well known major risk factors for oral cancer. For example, Döbróssy  reports The same pattern of estimates decreasing after adjustment for smoking and alcohol can be seen from the separate results for chewing tobacco, snuff and unspecific smokeless tobacco, although for snuff and particularly chewing tobacco numbers of estimates adjusted for both factors are very low. None of the analyses show any significant increase in risk following adjustment for both factors.2 value reduced from 94.9 to 71.6.The elevated random-effects estimate 5.15 for snuff based on unadjusted data  were for smokers or for nonsmokers. Eliminating these estimates so that all the data were for smokers and nonsmokers combined reduced the fixed-effect estimate only slightly, from 1.54 (1.40–1.69) to 1.53 (1.39–1.68) but increased the random-effects estimate from 1.87 (1.40–2.48) to 1.94 (1.43–2.65).Of the 38 estimates, five  were derived approximately assuming that there were no joint users of chewing tobacco and snuff. Replacing these by estimates for chewing tobacco in US studies and for snuff in Swedish studies increased the fixed-effect estimate to 1.62 (1.47–1.78) but left the random-effects estimate the same at 1.87 (1.40–2.50).2 = 1.15, p = 0.28). Although the number of published estimates is too few for reliable interpretation, this suggests that no bias to the null has been introduced by considering derived estimates.Of the 38 estimates used in the all smokeless tobacco analyses, only five were provided directly in the publications, the remaining 33 being derived from the data available. Although fixed-effect estimates were lower for the published data  than for the derived data , the difference was not statistically significant  of the total heterogeneity.Although 67.7% of the total heterogeneity could be accounted for by type of tobacco in the unadjusted analyses, the corresponding proportion for the smoking-adjusted risk estimates was only 15.1%. Although the smoking-adjusted effect estimates for chewing tobacco (1.29) and snuff (1.62) were elevated, this may reflect uncontrolled confounding by alcohol as no increase was seen in the analyses in Table Studies conducted in the US indicated an increased risk of oral cancer in both unadjusted data (by 76%) and adjusted data (by 39%). Though no significant effect was seen in Scandinavia or in other countries, numbers of estimates were relatively few , and heterogeneity by location was only demonstrated for the unadjusted data.Sex-specific overall risk estimates were higher for women than men. The fact that the sex difference was larger using unadjusted estimates suggests that at least part of it may be due to confounding.Hospital-based case-control studies provided higher risk estimates than those from other study designs, and their estimates were much more heterogeneous. The reason for this is unclear.Figure 2 = 63.1). Also shown in Table While the previous analyses are based on results, where available, for the whole population studied, including smokers and never smokers, six studies provided oral cancer relative risks specifically among never smokers. The estimates, two for snuff use and four for general smokeless tobacco use, are summarized in Table This report is based on data relating oral cancer risk to the consumption of chewing tobacco and snuff as used in Western countries. Our main analyses, based on results for smokers and nonsmokers combined where possible, considers 53 individual effect estimates based on 32 studies published from 1920 to 2005. This represents 11 more studies than considered in the previous most recent quantitative review of the evidence. 16,540 2 statistic for the 15 estimates was 94.9  by restricting attention to estimates adjusted for these two variables. Not only did such adjustment remove the substantial heterogeneity, but it removed the association, with an overall estimate for all smokeless tobacco use of 1.87 (1.40–2.48) reducing to 1.02 (0.82–1.28). The latter estimate is consistent with risk of oral cancer being independent of smokeless tobacco use, though a small increase cannot of course be totally excluded.Apart from the diversity of designs, samples, procedures, methods, and types of smokeless tobacco investigated, the individual effect estimates were themselves highly variable, particularly for snuff, where the IThe lower effect estimates seen in studies that adjust for alcohol and smoking need not wholly be due to the adjustment. For example, unadjusted estimates may derive from studies which are methodologically weaker, perhaps because smokeless tobacco was not a central issue. As such it is important to look at other information which may cast light on the role of confounding by alcohol or smoking. One way of attempting to eliminate potential confounding effects of smoking is to restrict attention to study of the effects of smokeless tobacco use in never smokers. Unfortunately only six studies provided effect estimates for oral cancer or data from which these could be calculated. Although the data are suggestive of an association, the meta-analysis estimates  which were adjusted at least for smoking. However, the corresponding estimates of 1.42 (0.99–2.03) and 1.28 (0.76–2.14) have quite low precision and provide little useful information on whether the associations truly differ. Anyway, they are likely to be biased upward by uncontrolled confounding by alcohol.For confounding to occur it is necessary for smokeless tobacco users and non-users to differ in their smoking and alcohol consumption. However relevant data seem quite limited and somewhat variable. The evidence seems more consistent for smoking where data indicating a positive relationship with smokeless tobacco use are seen in published studies in the USA ,71,72 anThe importance of confounding by alcohol consumption cannot, however, be fully resolved from the data available. Not only are the available data on the association of alcohol and smokeless tobacco very limited, they are unadjusted for age and race and usually for smoking or sex. Anyway the association may vary by type of smokeless tobacco, type of alcohol, country and period. It certainly seems to us that, at least in some studies, observed differences in risk of oral cancer associated with use of smokeless tobacco and type of tobacco used may be due to confounding by alcohol. In this context it should be noted that other potential sources of confounding are not even mentioned by most authors, like exposure to sunlight (being relevant for lip cancer) which might (as might the use of smokeless tobacco) show a social gradient due to an association with the type of occupation.Although conclusions based on adjusted estimates seem more convincing than those based on unadjusted ones, they are still not very reliable. One reason is the small number of available estimates. A second reason is the limited nature of the adjustment – thus many of the studies that adjusted for smoking did not take into account amount smoked e.g. ,74,75) a,75 a74,7A third reason for unreliability of the adjusted estimates is the variability of the findings. As the heterogeneity analysis revealed, study period was a major source, with smaller risks found in more recent studies. In fact, those studies which reported very large relative risk estimates for specific sites ,62,63 arOur analyses of smoking adjusted risks seemed to suggest that, compared to non-users, female users of smokeless tobacco might be at higher risk than male users, with fixed-effect relative risks of 2.51 vs 1.15. The result for females is, however, based on only two studies, of which only the Winn et al. study  found anOur analyses showed some indications of publication bias, a finding supported by others . SpecifiOur review is consistent with a general trend in appraising the potential risk of oral cancer from use of smokeless tobacco. In the mid 1980s smokeless tobacco was assessed as a risk factor for oral cancer by IARC and the US Surgeon General ,11. AlthThese reviews have led to discussion as to whether smokeless tobacco might be a less dangerous alternative to smoking for those who do not quit -89. UnliDetailed assessment of the overall risks and benefits of smokeless tobacco use to public health requires consideration of the whole spectrum of its possible health effects and is beyond the scope of this review. While it is clear that there are unique risks from smokeless tobacco use , there ES and RW work for Philip Morris International (PMI), R&D. Both receive their salary from PMI and both own shares in Altria, the holding company of PMI. PNL, founder of P.N. Lee Statistics and Computing Ltd., is an independent consultant in statistics and an advisor in the fields of epidemiology and toxicology to a number of tobacco, pharmaceutical and chemical companies.PNL conducted an unpublished review and meta-analysis on this subject in 2002. Although RW conducted and wrote up the present analyses, the earlier work provided an extremely helpful basis for comparison and validation at various stages. PNL also assisted with the preparation and layout of the final manuscript for publication. ES initiated the work and made substantial contributions to planning, analysis and writing. He continuously contributed to the work, revised the text critically, and also provided the final language corrections. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


Estrogens, environmental chemicals with carcinogenic potential, as well as oxidative and carbonyl stresses play a very important role in breast cancer (BC) genesis and progression. Therefore, polymorphisms of genes encoding enzymes involved in estrogen biosynthesis pathway and in the metabolic activation of pro-carcinogens to genotoxic intermediates, such as cytochrome P450C17α (CYP17), endogenous free-radical scavenging systems, such as glutathione S-transferase (GSTP1) and paraoxonase 1 (PON1), and anti-glycation defenses, such as glyoxalase I (GLO1), could influence individual susceptibility to BC. In the present case-control study, we investigated the possible association of CYP17 A1A2, GSTP1 ILE105VAL, PON1 Q192R or L55M, and GLO1 A111E polymorphisms with the risk of BC.The above-said five polymorphisms were characterized in 547 patients with BC and in 544 healthy controls by PCR/RFLP methods, using DNA from whole blood. To estimate the relative risks, Odds ratios and 95% confidence intervals were calculated using unconditional logistic regression after adjusting for the known risk factors for BC.CYP17 polymorphism had no major effect in BC proneness in the overall population. However, it modified the risk of BC for certain subgroups of patients. In particular, among premenopausal women with the A1A1 genotype, a protective effect of later age at menarche and parity was observed. As to GSTP1 and PON1 192 polymorphisms, the mutant Val and R alleles, respectively, were associated with a decreased risk of developing BC, while polymorphisms in PON1 55 and GLO1 were associated with an increased risk of this neoplasia. However, these findings, while nominally significant, did not withstand correction for multiple testing.Genetic polymorphisms in biotransformation enzymes CYP17, GSTP1, PON1 and GLO1 could be associated with the risk for BC. Although significances did not withstand correction for multiple testing, the results of our exploratory analysis warrant further studies on the above mentioned genes and BC. Breast cancer (BC) is both the prevailing malignancy and the most common cause of cancer death among women in Western countries . EstrogeMuch interest has long been addressed to cytochrome P450c17α (CYP17) and glutathione S-transferase 1 (GSTP1) genes, encoding enzymes involved in estrogen biosynthesis and metabolism  or in en2O2 (peroxidase-like activity), (c) preservation of HDL integrity and function and, finally, (d) prevention of LDL oxidation. Additionally, paraoxonase 1 is consistently implicated in the elimination of carcinogenic lipid-soluble radicals from lipid peroxidation [Moreover, the search for additional metabolizing polymorphic genes as potential susceptibility factors in BC, is needed. In particular, paraoxonase 1 gene (PON1), encodes a serum high-density lipoproteins (HDL)-associated enzyme , playingxidation . Therefoxidation . Two comxidation ,32. So fxidation ,34. It wGlyoxalase I (GLO1) gene encodes an anti-glycation defence enzyme that decreases the concentration of dicarbonyl compounds , such as methylglyoxal (MG), the most reactive glycation precursor . Since sWe would also like to highlight that all these four genes have been simultaneously considered in the present study because all related to the control of oxidative stress. In fact, they encode for enzymes that detoxify free radicals or reduce potential substrates for their production. As well known, there is evidence that oxidative stress, resulting from either an excess of reactive oxygen species (ROS) or a deficiency in antioxidant capabilities, may play a role in the etiology of BC ,37. In pThe study protocol followed the guidelines of our local ethics committee and the investigation was conducted with the ethical requirements defined in the Helsinki Declaration. All patients gave their informed consent to participate in the study.in situ (n = 46) or invasive (n = 292), were designated as "local". The age at diagnosis of BC was 55.3 ± 10.5 (SD) years and the patients had no previous history of BC or any other cancer. Control women (n = 544) were healthy Caucasian Italian women as well, randomly selected from the Umbrian Registry of Total Population, matched by age . The participation rate among controls was 87%. All these subjects were asked to complete a self-administered questionnaire on demographic factors, anthropometric measures, menstrual, reproductive, and breast feeding histories, use of contraceptives and exogenous hormones, medical and screening histories, first-degree family history of BC, occupational exposure, smoking history and alcohol consumption. All subjects were asked to provide a blood sample, to be used for genotyping. The median time between diagnosis and interview for BC cases was 2 months.Caucasian Italian women with a clinical and histological diagnosis of breast cancer (BC)  were enrolled in this study between June 2003 and July 2007 from the Breast Unit, Surgical Department of the University of Perugia, Umbria, Italy. Women with BC were identified via the Regional Cancer Registry, to which reporting of all malignant tumors is mandatory, and contacted via their doctors. In total, 547 women agreed to participate in the study, representing 89% of those who were contacted and found to be eligible. Histological staging of breast carcinoma was performed according to the current classification of the International Union Against Cancer . Women s2, 10 mM Tris-HCl pH 8.3, 50 mM KCl, and 1 U of Taq DNA polymerase. DNA amplification was performed conventionally; samples were subjected to 35 cycles of amplification consisting of 40 sec. denaturation, 30 sec. annealing, and 30 sec. extension. The final extension step at 72°C was extended to 10 min. The annealing temperature was optimized for each primer set.Genomic DNA was extracted from heparinized peripheral whole blood, using the QIAamp Blood Mini Kit . The nucleotide sequence of primers is shown in Table SNPs detection was based upon Restriction fragment length polymorphism (RFLP) analysis after PCR amplification. CYP17 gene maps to chromosome 10. A single T  to C  nucleotide substitution in the 5'-untranslated region of CYP17 creates a recognition site for the MspA1 restriction enzyme. GSTP1 gene has been mapped to a small region of chromosome 11. The region of genomic DNA flanking exon 5 of GSTP1 gene was amplified. The A to G substitution in GSTP1 leads to the creation of a recognition site for the Alw261 restriction enzyme. PON1 gene is located at chromosome 7. Single-base substitutions, which lead to a change of a glutamine residue into arginine at amino acid position 192, and a change of leucine to methionine affecting amino acid 55 of the paraoxonase protein encoded by human PON1, create recognition sites for Alw1 (PON1/192) and NlaIII (PON1/55), respectively. GLO1 gene maps to chromosome 6. A C to A substitution in GLO1 exon 4, which changes Ala111Glu in the encoded protein, leads to the loss of a recognition site for the SfaNI restriction enzyme. PCR products were, therefore, digested for 3 hr at 37°C using either MspA1, Alw261, Alw1, NlaIII, or SfaNI, respectively, and separated by electrophoresis in 12% polyacrylamide gels. The band patterns were developed by silver staining to identify the single base pair change. All genotyping was carried out by laboratory personnel blinded to case-control status of the samples, which included quality control samples for validation. Concordance for quality control samples was 100%.2 analysis, or by Fisher's exact test, where cell numbers were < 5. We used Bonferroni correction to account for multiple comparisons, and a two-tailed P value < 0.000862 was considered statistically significant.The obtained results were analyzed with the MedCalc statistical package . The associations between CYP17, GSTP1, PON1, GLO1 alleles and breast cancer (BC) were evaluated using unconditional logistic regression. All Odds ratio (ORs) and 95% confidence intervals (CIs) were adjusted for potential modifying factors, including age at menarche , age at first full-term pregnancy (FFTP) , history of benign breast disease , first-degree family history of breast cancer , waist-hip ratio (WHR) , use of oral contraceptive , alcohol consumption , smoking habits , level of education  and postmenopausal use of estrogens . Covariates as body mass index (BMI), breast cysts and physical activity were not included in the logistic model, because they did not alter the OR by > 5%, either separately or in combination. Since risk factors for BC significantly vary depending on menopausal status, we performed our analysis not only in the overall population, but also in premenopausal and postmenopausal subsets of women. Women who reported natural menopause or had undergone bilateral oophorectomy were classified as postmenopausal. Hysterectomized women with intact ovaries/ovary and women for whom the details of the operation were unknown were also classified postmenopausal, if they were no longer menstruating and were older than 50 years . All of the others were classified premenopausal. WHR, BMI, and age at FFTP were dichotomized on the basis of the median values for population controls. Estimates of statistical significance were calculated by standard χ2 test.Descriptive analysis included determination of standard deviation (SD) for cases and controls as well as Student's t-test to evaluate differences between means. A two sided probability value of less than 0.05 was considered to indicate statistical significance. For each group (controls and cases), allele frequencies were calculated by direct gene counting. Deviations from Hardy-Weinberg equilibrium (HWE) were tested by χThe main characteristics of the study subjects  and potential risk factors for breast cancer (BC) are shown in Table 2  and postmenopausal (79%) cases compared with the respective frequencies (66% and 65%) in the control subjects. Premenopausal women with the MM genotype had a significant increase for BC risk compared with the LL genotype. Postmenopausal women with the LM and MM genotype had a significant increase for BC risk compared with the LL genotype. PON1 Q192R and L55M genotypes were then analyzed according to tumor stage .The aetiological factors implicated in breast carcinogenesis are unclear, but estrogen levels , lifestyAs to GSTP1, we found that the major risk for BC was associated with individuals homozygotes for the Ile allele, while the Val allele appeared to be a protective factor against BC. A similar trend was presumably related with an activation of the enzyme at tissue level. Besides, experimental evidences showed that the Val variant may have either lower or higher specific activity and affinity depending on the substrate . On thisRegarding PON1, we found that women with PON1-192/QR and RR genotype had a lower risk of BC in comparison to women with the PON1-192/QQ genotype. This might be explained since, as expected, the Q to R substitution lead to the production of an enzyme with a higher detoxification activity against potentially carcinogenic products of oxidative stress and lipid peroxidation. This is supported by a quite recent study showing that individuals with a higher PON1 activity had a higher frequency of genotypes containing the R and L alleles ,49. WhenRegarding GLO1 polymorphism, our results indicate that the presence of the GLO1 E allele significantly increases BC risk. This might be explained by the fact that the A to E substitution, due to the SNP, may determine a conformational modification in the enzyme, leading to an isoenzyme with a lower detoxification capacity . As wellThe single SNP associations found in this study did not withstand correction for multiple testing by Bonferroni method. Additional data are therefore needed to corroborate our findings. These may be derived by replication in other genetic-epidemiological studies, preferably in combination with studies on the functional role of gene products that provide biological plausibility.We also determined the genotypic and allelic frequencies of CYP17, GSTP1, PON1-192 and-55 as well as GLO1 polymorphisms for each considered group (control and patients). When tested for HWE, the genotypic frequencies were in agreement with those predicted under HWE only for CYP17 polymorphism. Conversely, there were statistically significant deviations among the controls between the observed and expected genotypic frequencies for all the other polymorphisms. Such deviations, in literature, have been attributed primarily to a possibility of genotyping error . We haveIn conclusion, the results of this report suggest that polymorphisms of genes involved in the oxidative stress control may play a significant role in the development of BC. In particular, CYP17 polymorphism may modify the risk of BC only for certain subgroups, while GSTP1, PON1-192, PON1-55 and GLO1 polymorphisms appear to be common genetic traits likely associated with the risk of BC. However, the results of our exploratory analysis, while nominally significant, did not withstand correction for multiple testing. Therefore, further studies are needed to confirm our findings and to explore the exact molecular basis of our observations.BC: Breast cancer; CYP17: Cytochrome P450C17α; GSTP1: Glutathione S-transferase; PON1: Paraoxonase 1; GLO1: Glyoxalase 1; PCR/RFLP: Polymerase Chain Reaction/Restriction fragment length polymorphism; ROS: Reactive oxygen species; LDL: Low density lipoproteins; AGEs: Advanced glycation end-products; MG: Methylglyoxal; HDL: High density lipoproteins; SNP: Single nucleotide polymorphism; OR: Odds ratio; CI: Confidence intervals; FFTP: First full-term pregnancy; WHR: Waist-hip ratio; BMI: Body mass index; SD: Standard deviation.The authors declare that they have no competing interests.CA participated in the design of the study and its coordination. Performed the statistical analysis and drafted the manuscript. CDB carried out the burden of the molecular assays (PCRs and RFLP). VL and SG participated in the design of the study and in carrying out the initial phases of the molecular assays (DNA extraction). AR and FB participated in the design of the study and recruited all patients and collected all clinical data. VT conceived of the study 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:http://www.biomedcentral.com/1471-2407/9/115/prepub


TGFBR1*6A is a common hypomorphic variant of transforming growth factor β receptor 1 (TGFBR1). TGFBR1*6A is associated with an increased cancer risk, but the association of this polymorphism with osteosarcoma remains unknown. We have measured the frequency of TGFBR1*6A variants in osteosarcoma cases and controls.Our case-control study is based on 168 osteosarcoma patients and 168 age- and gender-matched controls. Blood samples were obtained and the TGFBR1*6A variant determined by PCR amplification and DNA sequencing. The odds ratio (OR) and 95% confidence interval (95% CI) for the TGFBR1*6A polymorphism were calculated by unconditional logistic regression, adjusted for both age and gender. Three models - dominant, additive and recessive - were used to analyze the contribution of the TGFBR1*6A variant to osteosarcoma susceptibility.Heterozygotic and homozygotic TGFBR1*6A variants represented 50.4% and 6.0% of the 168 cases, whereas the controls had 18. 5% and 1.3%, respectively. ORs for homozygosity and heterozygosity of the TGFBR1*6A allele were 4.6  and 2.9  in the additive model. There were significant increases in the TGFBR1*6A variants in osteosarcoma cases compared to control in all 3 models. Further analysis showed that TGFBR1*6A genotypes were not associated with gender, age, or tumor location. However, TGFBR1*6A was significantly associated with less metastasis.TGFBR1*6A, a dominant polymorphism of TGFBR1, is associated with increased susceptibility and metastasis spread of osteosarcoma. Osteosarcoma is the primary malignancy of bone that has peak occurrences in adolescence and after 50 years of age ,2. AlthoTGF-β signaling plays an important role in the development of tumors because it is a potent inhibitor of cell proliferation in most cell types ,5. TumorPatients with osteosarcoma admitted to the Department of Orthopedics at Tangdu Hospital from January 2001 to March 2009 were selected to participate in the study, making a total of 168 patients. The mean age in the cases was 30 years (range 10-71 years). All the cases were diagnosed without a familial cancer syndrome. As controls, DNA from 168 blood donors of mixed gender collected at Tangdu Hospital, Xi'an, China, were used. The controls were disease-free. Informed consent was obtained from all the subjects, and the study was approved by the ethical committee at Tangdu Hospital.TGFBR1*6A genotyping invovled the protocol of Song et al . PCR amp2 test. The Armigate treand test was used to calculate P for trends in the additive model. Results are presented as odds ratios (ORs) with 95% test-based confidence intervals (CIs). All P values were 2-sided and P < 0.05 was considered significant.Genotype distribution in osteosarcoma cases and controls was tested for the Hardy-Weinberg equilibrium, and the 2 variants were shown to be in equilibrium in both the patients and the controls. Differences in TGFBR1*6A variant carrier frequencies between cases and controls for the dominant and recessive models were assessed by the χThe frequencies of 9A/6A and 6A/6A genotypes were significantly different between the cases and controls. In 168 cases, 9A/9A, 9A/6A and 6A/6A genotypes re[resented 107, 51 and 10, respectively. In 168 controls, the genotypes of 9A/9A, 9A/6A and 6A/6A were 134, 31 and 3, respectively. Heteozygotic and homozygotic TGFBR1*6A variants were 50.4% and 6.0% in total 168 cases, while they were 18.45% and 1.29% respectively in total 168 controls. When the data were analyzed with the dominant, additive and recessive models, there were significant increases in TGFBR1*6A variants in osteosarcoma cases compared to control in all of the (P < 0.01). The result are shown in Table No significant statistic differences were shown in the association of TGFBR1*6A genotypes with clinical parameters between males and females. Nor was any significant difference found between patirnts under 20 and those over 20. The location of osteosarcomas in long tubular bones and the axial skeleton was not significantly associated with a particular mutation of the TGFBR1*6A genotype. However, in cases with distant metastasis, the frequency of TGFBR1*6A genotypes was significantly decreased compared with those without distant metastasis (P < 0.01). The result are shown in Table The 9A/6A and 6A/6A genotypes of TGFBR1 have a significantly higher frequency in osteosarcoma cases than in controls. TGFBR1*6A is a dominant susceptibility allele in the occurrence of osteosarcoma. The retrospective, non-randomized nature of most case-control studies limits the conclusion that can be reached from them. Because the subjects our study were from different regions of China for a long period from 2001 to 2009, the results have been more informative.The association between TGFBR1*6A and tumorigenesis seems to be different in our case. Previous reports indicated that TGFBR1*6A increased susceptibility of tumorigenesis and development in renal cell carcinoma, ovarian cancer, and breast cancer -14,16, aTGF-β signaling may not be the same in different tissues, which might explain the tissue-related manner of the association between TGFBR1*6A genotypes and cancer development. TGF-β signaling seems to be involved in tumorigenesis because it may behave aberrantly in its role as a potent inhibitor of cell proliferation in most epithelial and lymphoid cells . Tumors susceptibility to osteosarcoma.Our study firstly demonstrates the significant association between TGFBR1*6A and osteosarcoma, which may be due to the important role of TGF-β signaling in the control of osteosarcoma. Some data indicates that TGF-β can be an important factor in the occurrence of osteosarcoma. TGF-β receptor is expressed in osteosarcoma . TGF-β iOsteosarcoma has a high incidence in young children and is often located in long tubular bones . TherefoAnalysis of distant metastasis, however, showed that TGFBR1*6A is significantly related with metastasis in osteosarcoma. The decreased risk of distant metastasis of osteosarcoma in TGFBR1*6A variants suggests that TGF-β signaling is involved in the metastasis of osteosarcoma. Tumor metastasis is a complex process and many factors are involved in this process, TGF-β is only one of these factors. For example, TGF-β promotes tumor metastasis in cancer -36. AlthThis case-control study shows a significant statistical association between TGFBR1*6A variant and osteosarcoma in a Chinese population. The TGFBR1*6A variant is also significantly associated with the distant metastasis of osteosarcoma in the Chinese population studied.The authors declare that they have no competing interests.YSH and YP originally conceptualized the analysis, prepared the samples, and wrote the manuscript. WHL contributed to PCR and gene sequencing studies. YZ help with data analysis. BAM coordinated the study and participated in its design. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/10/169/prepub


Premature labor is a poorly understood condition. Estrogen is thought to play a key role and therefore the labor process may be affected by endocrine disruptors. We sought to determine whether or not an environmental toxicant, DDE, or dietary derived endocrine disruptors, daidzein and genistein, are associated with spontaneous preterm labor.Cases were defined as primiparous patients having a preterm delivery at or before 35 weeks following the spontaneous onset of labor. Controls were defined as primiparous women who delivered on the same day as the cases but at term gestation.Over approximately 1 year, 26 cases and 52 controls were recruited. Subjects agreed to have blood tests on day one postpartum for DDE and for the phytoestrogens genistein and daidzein.The mean concentration of DDE was similar in the case and control groups: 4.29 vs 4.32 ng/g lipid p = .85. In the case group, 13/26 had detectable levels of daidzein (range 0.20 – 1.56 ng/ml) compared to 25/52 controls (range 0.21 – 3.26 ng/ml). The mean concentration of daidzein was similar in cases compared to controls: 0.30 vs .34 ng/ml p = 0.91. Of the case group,14/26 had detectable levels of genistein (range 0.20 – 2.19 ng/ml) compared to 32/52 controls (range 0.21 – 2.55 ng/ml). The mean concentration of genistein was similar in cases compared to controls: 0.39 vs 0.31 ng/ml, p = 0.61.The serum levels of DDE in this population were found to be low.There appears to be no relationship between serum concentrations of DDE, daidzein, and genistein and spontaneous preterm labor in our population. The inability to identify an effect may be related to the comparatively low concentrations of DDE in our population and the rapid and variable reduction of phytoestrogens from women in labor. Premature delivery occurs after spontaneous labor in approximately 6% of pregnancies (<37 weeks gestation) and remains the single, most important, cause of perinatal mortality and morbidity ,2. The pThe phytoestrogens are of increasing interest owing to their estrogenic properties and ubiquitous exposure from a variety of foods and the ability of individuals to alter the exposure through soy supplementation . These cDDT and PCBs have also been of interest in the pathogenesis of premature labor and have shown to increase myometrial contractility in-vitro -13. A laCases were defined as primiparous women having a premature delivery at greater than 23 weeks and less than 35 weeks gestation following spontaneous labor. The gestational age of <35 weeks was chosen as this corresponds to clinically significant premature labor and this definition has been used in many contemporary studies in prematurity -23. CaseThe demographic characteristics of the subjects were similar and are provided in Table In the case group, 13/26 had detectable levels of daidzein (range 0.20 – 1.56 ng/ml) compared to 25/52 controls (range 0.21 – 3.26 ng/ml). Figure Our study did not demonstrate an association between spontaneous premature labor and serum DDE levels. Our results, considered with the results of another case control study on a contemporary population, would sAlthough our previous work had suggested that the placenta and uterus are being exposed to significant levels of the phytoestrogens daidzein and genistein, their levels were not associated with spontaneous premature delivery. One limitation is that we were not able to measure a cumulative exposure to the phytoestrogens and that we had to rely on a single post-partum serum level. It would be difficult and expensive to have performed serial measurements and, as it is difficult to predict the onset of premature labor, even a prospective study would have similar limitations. Genistein and diadzein levels appear to peak approximately 6 hours after intake and their half lives are approximately 6 and 8 hours respectively. InevitaLike any observational study confounding bias is always a possible explanation for the observed results. We controlled for the two strongest potential confounding factors: primiparity and previous preterm delivery by restriction. Maternal age was controlled for by matching. We also adjusted for other confounders such as maternal weight and smoking in the analysis. We did not however have information on socioeconomic status, a modestly strong risk factor and there is always the possibility that unknown confounders could have affected our results.As with any negative study, a possible explanation for our results is inadequate power. While we accept this potential criticism, we point out that not only were the differences we observed not statistically significant but that our point estimates were also very close to the null value. However, it is still possible that the effect of these phytoestrogens may be mediated by complex interactions with other estrogenic compounds, as has been indicated in studies using the rat uterotrophic assay . As our In this case control study, maternal serum levels of DDE, and the phytoestrogens, daidzein and genistein were not associated with spontaneous premature delivery. There is still a possibility that higher exposure to these substances may contribute to premature labour.DDE ethylene)DDT ethane)HCG Human Chorionic GonadotropinPCB Polychlorinated BiphenylsThe author(s) declare that they have no competing interests.All authors contributed to developing the original protocol. CS recruited the patients. SC supervised the analysis of the serum samples. SW performed the analysis. All the authors contributed to and approved the final manuscript.


AccuProbe culture identification kits for the Mycobacterium tuberculosis complex, M. avium complex and M. avium were used to identify the isolates. Spoligotyping and Insertion Sequence (IS) 1311 and IS1245 Restriction Fragment Length Polymorphism analysis (RFLP) were used to further characterise the isolates.Bovine tuberculosis is a zoonotic problem in pastoral cattle and communities in Uganda. Tuberculin tests in pastoral cattle had shown a high herd but low animal prevalence, with a high proportion of avian reactors. No work had been done to identify the mycobacterial species involved. The objective of the study was to isolate and characterise Mycobacterial species causing tuberculous lesions in slaughtered animals. Lesioned organs compatible with bovine tuberculosis in slaughtered cattle from pastoral areas in Uganda were collected and cultured to isolate M. bovis, 3 as M. avium subsp.hominissuis, 1 as M. intracellulare, 1 as a mixed culture of M. bovis and M. avium sp. and 1 as M. avium sp. and unidentified mycobacteria. Eleven other mycobacteria outside the tuberculosis and avium complex groups were also isolated. Ten new spoligopatterns grouped into three clusters were identified from M. bovis isolates. Two of the three M. avium subsp.hominissuis isolates showed similar patterns on the IS1311 RFLP but all were different on the IS1245 RFLP.Of the 61 lesioned organs and tissues cultured, 19 isolates were identified as M. bovis confirms the ongoing infection with spoligotypes unique to Uganda. Isolation of environmental mycobacteria could explain the high avian or non specific tuberculin reactor patterns commonly observed in pastoral cattle and suggests their pathogenic or opportunistic role in the infection of cattle with disseminated bovine tuberculous lesions.The isolation of  Mycobacterium bovis infections are of major importance in many developed and developing countries, including Uganda  was added. After vortexing, the specimens were left at room temperature for 15 min. Subsequently 5 ml of sterile 0.067 M phosphate buffer (pH 6.8) was added and the mixture was centrifuged at 3660 × g for 15 min. The supernatant was discarded and the sediment was inoculated on Lowenstein-Jensen  media with and without pyruvate (0.6%) and incubated at 37°C. Tubes were read weekly for up to 12 weeks. Of the 61 lesioned specimens collected and cultured, 37 grew ® INC. San Diego, California) as described by the producer. Results were considered positive when Relative Light Units (RLU) were greater than 30 000, repeat range; 20 000–29 000 and negative; below 20 000. Samples negative on the MTC kit were examined further on the AccuProbe MAC (M. avium complex) and AccuProbe M. avium culture identification kits. Final results were interpreted as follows; cultures positive on the M. avium and MAC culture identification kits were considered M. avium species and cultures negative on the M. avium kit and positive on the MAC kit were identified as M. intracellulare. Samples negative on all AccuProbe culture identification kits were grouped as unidentified mycobacteria.All acid fast bacteria, determined by the Ziehl-Neelsen (ZN) staining technique, were examined by the AccuProbe MTC (mycobacterium tuberculosis complex) identification kit . Normalisation of the fingerprints was done using the molecular weight standard 1 kb DNALadder . The BioNumerics software was used to calculate Dice coefficients of similarity and to cluster the isolates and generate dendograms by the unweighted-pair group method using linkage averages. The optimisation and tolerance settings were both set at2.00%.M. avium sp. were further examined using IS1245 and IS1311 Restriction Fragment Length Polymorphism (RFLP). The DNA extraction, RFLP analysis and interpretation of results were performed as previously described [1245 and IS1311 RFLP.Cultures identified as escribed , using tThe author(s) declares that there are no competing interests.mycobacteria, molecular studies and drafted manuscript. R.K: Conception and design of the study, general conduct of the study, acquisition of data, isolation and culture of mycobacteria. AL: Participated in the conception and design of the study, acquisition of funding, general conduct of the study, isolation and culture of mycobacteria and critical revision of the manuscript for important intellectual content. J.O-A: Participated in the conception and design of the study, general supervision of the research group in Uganda. General conduct of the study, acquisition of data, isolation and culture of mycobacteria, molecular studies and critical revision of the manuscript for important intellectual content. ES:Acquisition of funding, general supervision of the research group in Norway, participated in the conception and design of the study, acquisition of data, drafted the manuscript. B.D: Participated in the design of the study, general supervision of the laboratoty work in Norway, isolation and culture, molecular studies and critical revision of the manuscript for important intellectual content. TBJ: Carried out the molecular genetic studies, participated in the sequence alignment and drafted the manuscript. BD:Isolation and culture, molecular genetic studies and critical revision of the manuscript for important intellectual content.JO: Principal investigator, participated in the conception and design of the study, general conduct of the study, acquisition of data, isolation and culture of All the authors have read and approved the final version of the manuscript.


The liver function test (LFT) is among the most commonly used clinical investigations to assess hepatic function, severity of liver diseases and the effect of therapies, as well as to detect drug-induced liver injury (DILI).To determine the relative contribution of genetic and environmental factors as well as test and quantify the effects of sex, age, BMI and alcohol consumption to variation in liver function test proteins - including alanine amino transaminase (ALT), Albumin, gamma glutamyl transpeptidase (GGT), total bilirubin, total protein, total globulin, aspartate transaminase (AST), and alkaline phosphotase (ALP) - using the classical twin model.Blood samples were collected from a total of 5380 twin pairs from the TwinsUK registry. We measured the expression levels of major proteins associated with the LFT, calculated BMI from measured weight and height and questionnaires were completed for alcohol consumption by the twins. The relative contribution of genetic and environmental factors to variation in the LFT proteins was assessed and quantified using a variance components model fitting approach.2 ranging from 20% to 77%); (2) other than GGT, the LFTs are all affected to some extent by common environmental factors (c2 ranging from 24% to 54%); and (3) a small but significant proportion of the variation in the LFTs was due to confounding effects of age, sex, BMI, and alcohol use.Our results show that (1) variation in all the LFTs has a significant heritable basis (hVariation in the LFT proteins is under significant genetic and common environmental control although sex, alcohol use, age and BMI also contribute significantly to inter-individual variation in the LFT proteins. Understanding the underlying genetic contribution of liver function tests may help the interpretation of their results and explain wide variation among individuals. The liver function test (LFT) is a routinely ordered clinical investigation used to assess hepatic function, disease severity In the clinic, LFT results are compared to and interpreted according to reference charts containing internationally accepted normal-levels for these proteins. However many patients are investigated for having levels outside the normal range without any obvious pathology or responsible factor being detected. The most common biomarkers that are routinely measured as part of the LFT include proteins which are specifically produced by the liver – such as ALT and albumin (ALB) – as well as those that are more ubiquitously produced, for example AST, which is also produced by red blood cells and cardiac/ skeletal muscles, and ALP which is also produced in bones and kidneys.At the population level, previous studies have shown that some of the LFT proteins are highly variable, affected by various epidemiological factors including body mass index (BMI) Twin studies provide one of the best epidemiological designs to gain insights into the contribution of genetic and environmental factors to population variation in clinical or evolutionary traits. Here, we use the classical twin design to investigate the relative contributions of genetic and environmental factors to variation in the most common liver biomarkers including alanine amino transaminase (ALT), albumin (ALB), gamma glutamyl transpeptidase (GGT), total bilirubin (TBILIRB), total protein (TPROT), total globulin (TGLOB), aspartate transaminase (AST) and alkaline phosphotase (ALP). Furthermore, we directly test for the effect of age, sex, BMI, and alcohol use on variation in the LFT proteins.The study was approved by St. Thomas' Hospital Research Ethics Committee (EC96/439 Twins UK) and all participants provided written informed consent.2)). Summary characteristics of the cohort and the measured proteins are presented in A total of 5380 twins, comprising of 1804 DZ pairs and 886 MZ pairs , from the TwinsUK database at St. Thomas' Hospital All assays were performed on a Synchron LX20 automated multi channel analyzer . ALT activity was measured by a kinetic rate method (within run precision 4.3 and between run 4.6 at 20 iu/L). Serum albumin concentration was measured by means of a bichromatic digital endpoint methodology using bromcresol purple (BCP) reagent (within run precision 0.2 and between run 1.3 at 48 g/L). GGT activity was measured by an enzymatic rate method (within run precision 3.5 and between run 5.3 at 86 iu/L). BILIRB was measured by a timed endpoint Diazo method (within run precision 3.0 and between run 4.5). TPROT concentration was measured by means of a rate biuret method (within run precision 0.7 and between run 0.87 at 75 g/L). AST activity was measured by an enzymatic rate method (within run precision 0.7 and between run 1.0 at 175 iu/L). ALP activity was measured by a kinetic rate method using a 2-amino-2-methyl-1-propanol (AMP) buffer (within run precision 1.8 and between run 2.1 at 150 iu/L).Twin studies are the optimum epidemiological design to study and partition population variation of a trait into genetic and non-genetic - shared and unique environmental - components. The classical twin design assumes that monozygotic twins share 100% of their genes and shared environment whereas dizygotic twins share on average 50% of their genes and 100% of shared environment. Thus any greater similarity between MZ as compared to DZ twin pairs is attributed to genetic factors 10 transformed, ALT, BILIRB, TPROT and AST were square root transformed, and ALB was normally distributed. Covariates age, BMI and alcohol consumption were all continuous variables as well.All measures of liver function test proteins were recorded as continuous variables. However, many were not normally distributed so appropriate statistical transformations were carried out prior to analyses; Total protein, GGT, TGLOB and ALP were logDemographic differences between MZ and DZ twins were assessed by comparing means, medians and standard deviations . We carrOrdinary least squares regression was used for analysis of covariates, age, BMI and sex. For alcohol consumption, truncated Gaussian regression was used to assess the independent contribution of alcohol consumption to variation of protein assay. Relatedness between the twins was accounted for using the “cluster” option in STATA. All preliminary analyses were performed using STATA 10 2) can be defined as the proportion of the total variance attributable to additive genetic variation or h2 = A/P. The significance of the variance components A, C, and D are assessed by comparing the model fit between a full model and a nested model in which each term is sequentially set to zero. Model fit is assessed using the likelihood ratio test. The purpose of the model fitting procedure is to explain the pattern of observed variances and covariances as accurately as possible. Genetic modeling was carried out using the computer package Mx, which has been specifically designed for analysis of twin and family data Standard methods of quantitative genetic analysis were used to model latent genetic and environmental factors influencing sibling covariance for MZ and DZ twins. Genetic model fitting  was used for the decomposition of the observed phenotypic variance (P) into additive (A) & dominant (D) genetic components and environmental components shared by both twins (C) & unique to each twin (E). The latter also includes measurement error. Dividing each of these components by the total variance yields the different standardized components of variance. For example the (narrow sense) heritability , although explicitly modeling the different MZ/DZ means and variances did not qualitatively affect the results in this case (data not shown).The mean and standard deviations of the LFT proteins, BMI, alcohol consumption, and age range for the MZ/DZ twin populations within each sex group were comparable (p>0.05). Significant differences by zygosity were noted for ALT in the male twin cohort . As expected we observed a significant difference in BMI  and alcohol consumption  between the male and female twins. Overall, male twins had a higher BMI and consumed significantly more alcohol than female twins.Age was comparable between male and female twin pairs (χ−4) with the proportion of variability explained (R2) ranging from 1% in the case of ALP to 6% for GGT and BILIR .As the mean and range of age was comparable across the male/female sex strata we report results from pooled analyses . Age was−4, R2 ranging from 2% for ALT in females to 11% for ALT in males), and ALB, BILIRB and ALP in female twin cohort only .BMI was significantly associated with ALT and GGT, in both the male and female twin cohort . In the female twin cohort alcohol consumption was significantly associated with GGT, TGLOB and ALP (p<0.002).The results of genetic model fitting are shown in In this study we used the classical twin design to gain insight into the relative contribution of genetic, environmental, as well as confounding effects of sex, age, body mass index and relative alcohol use to variation in the all major liver function test proteins.Our results show that (1) in agreement with previous studies, variation in all the LFT proteins is highly variable; (2) a significant proportion of the variation (up to 11%) in the LFTs was due to confounding effects of age, sex, BMI, and alcohol use; (3) variation in all LFTs has a significant heritable basis; and (4) other than GGT, the LFTs are all affected to some extent by common environmental factors.Our results are in general agreement with previous cited studies that have investigated the effects of age Except total protein and total globulin, we observed significant sex differences with regards to variation in all the LFTs. For instance, our results show that up to 6% of the variation in bilirubin levels were due to sex. Furthermore, sex stratified analysis showed that BMI contributes significantly and differentially to the variation in many of the LFTs in our study. For example, up to 8% of the variation in ALP in females is due to variation in BMI whereas BMI contributes insignificantly to ALP variation in males. Moreover, although BMI is significantly associated to variation in ALT in both males and females the explained proportion of the variation in ALT by BMI is hugely different across sex groups; 2% in females and 11% in males. Finally, alcohol consumption was significantly associated with GGT, ALP, and total globulin in females. Whilst in the males, we showed that there was a trend of association with all the LFT proteins except ALT and ALP largely due to the fact that the male cohort was approximately 10 fold smaller than the female cohort. Interestingly, ALT, GGT and ALP were found to be significantly affected by all four confounding variable investigated - age, BMI, sex, alcohol use - highlighting the present dilemma and difficulty faced by clinicians in interpreting liver function test results.Our study represents the largest and most extensive study to investigate the influences of genes and environment on the major LFT proteins. Our data shows that, quantitative variation in all the major LFT proteins – corrected for the effects of age, sex, BMI, and alcohol use – is under strong genetic control with narrow-sense heritability ranging from 15 to 77%. Minor differences were noted in the choice of the best fitting model for ALT, ALB, BILIRB and ALP, when we carried out sex stratified analyses, although point estimates of the A and C variance components were very similar (data not shown). We have also performed all our analysis by excluding male twins and our results were unchanged.Our study is in general agreement with two previous studies. A small-scale study from the Danish twin registry consisting of 290 elderly twin pairs (age range between 73–102 years) were measured for a subset of the LFTs by Bathum et al 2 = 0.5, p<1×10−5) as well as ALT and AST  although GGT and AST levels were not correlated .A larger study by Whitfield et al. A major, previously unreported finding in our study is the strong influence of common environmental factors underpinning variation in the LFTs. Our result showed that variation in all but one of the LFT proteins (GGT) are strongly influenced by common environmental factors. In fact, in most cases the effect of common environment is either comparable or larger than the heritability of the trait. Although commonly highlighted as a strong motivation for undertaking a twin rather than a family based study, common environmental factors are seldom reported together with genetic factors in the same study. This is in large part due to lack of power in most twin studies to dissect the total familial component into its sub-components of genetic and common environments Although our study represents the largest and most extensive genetic epidemiological study of the LFT protein, there are a few potential limitations, which warrant further discussion. First, our study was conducted in twins, which have been previously shown to be representative of population-based samples for a wide range of common medical conditions and lifestyle characteristics Second, although all the participant twins were deemed generally healthy – i.e. not suffering from any chronic diseases or taking prescribed medication - our results may have been influenced to an extent by unforeseen perturbations due to an infection, early stages of a disease, or a short course of an over the counter medication or dietary supplement Third, due to non-technical reasons AST was only measured in a sub-sample of the female twin cohort (98MZ & 477DZ) and not in males at all. We are currently looking into increasing our sample size in the case of AST for future genetic association studies.Fourth, due to historical reasons the TwinsUK database is predominantly female cohort and so we only had access to a limited number of male samples (no more than 500 twins).Fifth, although not a major component of our study, the quality of data on alcohol consumption was not optimal. First, it was “recall” data collected from self-reported questionnaires and second our alcohol data is not representative of the drinking status of the individuals at the time at which the blood samples were taken for biochemical assays of the LFTs. For example, previously it has been reported that the amount of alcohol consumed in the past 30 days is an important indicator of high levels of GGT Finally, our conclusions are restricted to Northern European Caucasian populations and the results need to be replicated in other independent cohorts of different ancestry.In conclusion, our study suggests the presence of sex differences and effect of age, BMI and alcohol consumption in variation of liver function proteins among individuals. Substantial heritability was observed for all the LFTs with the quantitative genetic modeling showing that a great proportion of the variation in liver function proteins are due to additive genetic effects stimulating further research to identify the responsible genes/region. Identifying the genes that influence the widely used liver function test will be very important in the future interpretation of their results and possibly reduce the need to further investigate subjects with high levels due to normal genetic variation.


Markers of crypt cell proliferation are frequently employed in studies of the impact of genetic and exogenous factors on human colonic physiology. Human studies often rely on the assessment of tissue acquired at endoscopy. Modulation of cell proliferation by bowel preparation with oral laxatives may confound the findings of such studies, but there is little data on the impact of commonly used bowel preparations on markers of cell proliferation.Crypt length, crypt cellularity and crypt cell proliferation were assessed in biopsies acquired after preparation with either Klean-Prep or Picolax. Crypt cell proliferation was assessed by whole-mount mitotic figure count, and by two different immunohistochemical (IHC) labelling methods (Ki-67 and pHH3). Subsequent biopsies were obtained from the same patients without bowel preparation and similarly assessed. Parameters were compared between groups using analysis of variance and paired t-tests.There were significant differences in labelling indices (LI) between biopsies taken after Klean-prep and those taken after Picolax preparation, for both Ki67 (p = 0.019) and pHH3 (p = 0.017). A similar trend was seen for whole-mount mitotic figure counts. Suppression or elevation of proliferation parameters by bowel preparation may mask any effect due to an intervention or disease.Commonly used bowel preparations may have significant and different effects on crypt cell proliferation. This should be taken into account when designing studies and when considering the findings of existing studies. Although the risk of significant fluid or electrolytes shifts is reduced with PEG based preparations[Oral laxatives are routinely used to cleanse the colon prior to colonoscopy, to permit clear visualisation of the colonic mucosa and minimise the risk of infection. Hyperosmotic preparations such as sodium phosphate (NaP) and Picolax motility. The eff motility. Klean-Pparations, the larparations.Ideally, a bowel preparation agent should have minimal impact on the colonic mucosa. However, endoscopic and histological abnormalities have been reported, particularly with NaP, and include inflammation, haemorrhage, focal active colitis (FAC), erosion of the surface epithelium and apthoid ulceration -7. One sCell proliferation is a commonly used end-point in studies of the effects of exogenous and genetic factors on colonic mucosal function and cancer risk. However, data on the impact of bowel preparation agents on cell proliferation is sparse. A 136% increase in colonocyte Ki-67 labelling index has been reported in patients prepared with NaP, when compared to the same patients, unprepared. Sodium To date, there are no published studies examining or comparing the direct effects of sodium picosulfate and PEG on colonocyte proliferation. This study aims to compare the effects of these two commonly used bowel preparative agents on immunohistochemical and whole-mount derived indices of cell proliferation.Patients attending endoscopy clinics at the Northern General Hospital, Sheffield for an initial colonoscopy were recruited to the study. Patients were all male, over 40 and had normal bowel health. Smokers and diabetics were excluded. Patients prepared for endoscopy with Klean-Prep (n = 4) or Picolax (n = 3) were invited to return four weeks later for sigmoidoscopy without bowel preparation. Two biopsies were taken at the recto-sigmoid junction. This study was approved by the North Sheffield Research Ethics Committee (Reference number: 06/Q2308/93). Informed consent was obtained from all patients.The range of markers currently utilised for the assessment of colonocyte proliferation is wide, and may in part account for the variability seen between studies. In this study we employed two recognised immunohistochemical markers of cell proliferation, Ki-67 and pHH3. Ki-67 is expressed throughout the cell cycle, whilst pHH3 (phosphorylated histone H3) is detectable in increasing quantity during the progression from from interphase to prophase. This correlates with a progression from granular to diffuse immunohistochemical staining patterns. Counts Biopsies were fixed for 24 hours in formalin before paraffin embedding and cutting of serial 4-micron sections at 40-micron intervals. Endogenous peroxidise activity was blocked with 2% hydrogen peroxide. Heat-induced eiptope retrieval (HIER) was performed using a sodium citrate buffer (pH 6.0) for Ki-67 and EDTA (pH 8.0) for pHH3. Non-specific immunoglobulin binding sites were blocked with normal horse serum. Slides were incubated for 1 hour with primary antibodies to Ki-67 (Vector Laboratories VP-K 452) and pHH3 (Upstate 06–570) at 1:100 and 1:200, respectively. Staining was visualised with the Vectastain Universal Elite kit and DAB peroxidise substrate (Vector Laboratories). Sections were counter-stained with Gill's haematoxylin.A maximum of 10 well-orientated hemi-crypts per biopsy, showing the entire length of the crypt wall from the base abutting the muscularis mucosa through to the junction with the surface epithelium, were included. Images were captured at 20× magnification with a Nikon D5-M camera at 2560 × 1920 resolution, stored without compression and analysed using Nikon NIS-Elements D (v 2.30) software. All cell counts and measurements were made by one observer blind to the status of the biopsy. A subset of scores was confirmed by a second independent observer.The number of cells showing positive nuclear staining for each antigen was recorded as a fraction of the total cell count per hemi-crypt (labelling index:LI). For pHH3, the LIs for granular and diffuse staining were recorded separately. Crypt lengths for up to five crypts per biopsy were determined.Whole biopsies were fixed in Carnoy's fluid, then stored in 70% ethanol. Rehydration through 50% and 25% alcohols and PBS was followed by hydrolysis in 5 M NaOH at room temperature. After staining with Schiff's reagent for 90 minutes, the biopsies were transferred to 45% acetic acid for dissection. Small groups of crypts were isolated from surrounding tissue under a dissecting microscope, then transferred to slides and squashed with thumb pressure under a cover-slip with aqueous mountant. Mitoses were counted at ×40 in up to ten whole crypts per biopsy, through the full crypt depth. Nuclei in all phases of mitosis were counted and expressed as the number of mitoses per crypt.Statistical analysis was carried out with SPSS (version 11.0). The distribution of values for each parameter was tested for normality using the Kolmorogov-Smirnov test. Indices for paired biopsies  were compared using the paired Student t-test. One-way ANOVA was used to test for the equality of the means for each parameter between each group , and the unpaired t-test was used to test for differences between biopsies taken after Klean-prep and those taken after Picolax.Immunohistochemical methods for the assessment of cell proliferation in small biopsies suffer from a number of potential disadvantages. Firstly, it can be difficult to obtain high yields of assessable crypts in histological sections. Secondly, there is concern over the impact of the so called "denominator effect" on the LI. Counting mitotic figures in whole-mount preparations avoids these issues, but does suffer the disadvantage that a second biopsy must be obtained if conventional histolological examination  is required. We compared the statistical reliability of both approaches using the Cronbach's alpha statistic -a measure of between and within-case variability, ranging from 0 to 1 with increasing reliability. The scores were 0.9476 and 0.9164 for Ki-67 and whole mount mitotic counts, respectively, indicating that robust measurements can be obtained without exhaustive sampling of crypts. Moreover, scores from both methods correlated well . There is also concern that individual antibody targets may not display expression that is truly restricted to proliferating cells, and therefore give potentially spurious results. It is reassuring that the immunohistochemical markers used in this study both showed similar results.Two immunohistochemical markers of cell proliferation, Ki-67 and pHH3, were measured. Colonocyte Ki-67 LI decreased after preparation with Klean-Prep, when compared with unprepared bowel from the same patients Figure . DespitePicolax had an opposing effect on colonocyte proliferation. Ki-67 LI increased after preparation with Picolax, although this effect was not significant  and the unavailability of paired biopsies for all markers tested. All data, paired or otherwise, was therefore redistributed into treatment groups and subjected to analysis of variance (ANOVA) testing. One-way ANOVA revealed significant differences between groups , demonstrating, as for the paired analysis, a decrease in proliferation with Klean-Prep, and an increase with Picolax, when compared to unprepared bowel , or between groups by one-way ANOVA (p = 0.437). Similarly, there was no significant difference in crypt length between pairs , or between groups by one-way ANOVA (p = 0.0786) (data not shown). The cellular homeostasis observed despite increased proliferation could be attributed to two factors. Firstly, the interval between bowel preparation and biopsy  may have been insufficient for any cells entering cell division to complete the cycle. Secondly, the observed homeostasis may be attributable to increased apoptosis. Our assessment of apoptotic indices recorded very low rates of background apoptosis in all samples, and showed no alteration in response to bowel preparation (data not shown).A randomisation bias on clinical grounds cannot be ruled out in the allocation of patients to Klean-Prep or Picolax preparation; this, along with the small sample size, may in part account for the lack of statistical significance seen between prepared patients and unprepared controls. Similarly, whilst patients prepared with Klean-Prep or Picolax fasted prior to colonoscopy, control patients did not, and potential confounding influences of luminal nutrition and luminal workload on proliferation indices should be noted. However, our analyses, and the central finding of this study, show an opposing direction of change between Klean-Prep and Picolax that is consistent for all proliferation measures, and cannot be a feature of starvation. Although the small number of cases and the single sampling site inevitably means that the results should be interpreted with caution, this pilot study has informed our choice of procedural homogeneity for a separate, larger study requiring repeat endoscopies.This study demonstrates that Picolax and Klean-Prep have significantly different effects on crypt cell proliferation. Suppression or elevation of proliferation parameters by bowel preparation may mask any effect due to an intervention or disease.These data underscore the importance of ensuring that all patients participating in studies in which crypt proliferation is an outcome measure are treated with identical bowel preparation. Bowel preparations used in such studies should be reported, and the data interpreted with caution.The authors declare that they have no competing interests.LJC carried out the analyses and was responsible for drafting the manuscript. JPB contributed to the analyses, performed the statistical analysis and helped to draft the manuscript. EAW participated in the design of the study and assisted with patient recruitment. SR was responsible for patient recruitment and collected biopsy material for the study. BMC conceived the study, participated in its design 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:


Angiotensin converting enzyme (ACE) is a possible candidate gene that may influence both body fatness and blood pressure. Although several genetic studies have been conducted in adults, relatively few studies have examined the contribution of potential candidate genes, and specifically ACE I/D, on adiposity and BP phenotypes in childhood. Such studies may prove insightful for the development of the obesity-hypertension phenotype early in life. The purpose of this study was to examine differences in body fatness and resting blood pressure (BP) by ACE I/D genotype, and determine if the association between adiposity and BP varies by ACE I/D genotype in children.152 children  were assessed for body composition (% body fat) using dual energy x-ray absorbtiometry and resting BP according to American Heart Association recommendations. Buccal cell samples were genotyped using newly developed PCR-RFLP tests for two SNPs (rs4341 and rs4343) in complete linkage disequilibrium with the ACE I/D polymorphism. Partial correlations were computed to assess the ociations between % body fat and BP in the total sample and by genotype. ANCOVA was used to examine differences in resting BP by ACE I/D genotype and fatness groups.Approximately 39% of youth were overfat based on % body fat . Body mass, body mass index, and fat-free mass were significantly higher in the ACE D-carriers compared to the II group (p < 0.05). BP was not significantly different by ACE I/D genotypes. In the total sample, correlations between adiposity and BP ranged from 0.30 to 0.46, and were not significantly different between genotypes. When grouped by genotype and body fat category, the overfat D-carrier subjects had significantly higher SBP and MAP compared to the normal fat D-carrier and normal fat II groups (p < 0.05).ACE D-carriers are heavier than ACE II children; however, BP did not differ by ACE I/D genotype but was adversely influenced in the overfat D-carriers. Further studies are warranted to investigate the genetics of fatness and BP phenotypes in children. Currently, there is considerable interest in the genetics of complex human phenotypes such as obesity and hypertension. The estimated heritability of adiposity measures range from 25–40%  and seveAngiotensin converting enzyme (ACE) is a possible candidate gene that may have pleiotropic effects. ACE is an important regulatory enzyme of the renin-angiotensin-aldosterone (RAA) system, which is a complex system that plays a critical role in maintaining blood pressure homeostasis. ACE converts inactive angiotensin I into active angiotensin II (vasoconstrictor) and inactivates bradykinin and kallidin (vasodilators). Thus, activation of the RAA system results in a vasopressor response mainly through the actions of ACE. The ACE gene is located at 17q23 and contains a polymorphism distinguished by either an insertion (I) or deletion (D) of a 287 base pair segment in intron 16. The ACE DD genotype has been associated with higher levels of ACE , BP 3,6. AlthougIt is well-known that a positive relationship exists between several measures of body size and adiposity  and BP across the lifespan . SimilarGiven the paucity of studies in this emerging area of research, the purpose of this study was to twofold: 1) examine the differences in adiposity and BP between ACE I/D genotypes among young children, and 2) examine if the association between adiposity and BP was modified by the ACE I/D genotype.The participants in the study were young children  ages 3–12 years from a rural U.S. Midwestern community . The participants were recruited through written and/or verbal advertisements. Parental consent was obtained from all participants, and the procedures were also explained to the child participants. The study protocol was approved by the University of Nebraska at Kearney Institutional Review Board.2.Stature and body mass were measured according to standard procedures . StatureWhole-body DXA scans were performed with the subject in light clothing while lying supine using a Lunar DPX-L densitometer . Percent body fat (%BF), fat mass (FM) and fat-free mass (FFM) were determined using the pediatric medium scan mode (software version 1.5d). The Lunar DPX-L densitometer has previously been cross-validated using the pig carcass in the pediatric weight range . ReliabiIn addition to using the continuous variable of fatness in the statistical analysis, children were classified into one of two groups  based on the cut-points for % body fat used in the FITNESSGRAM test assessment program  as deterResting BP was measured in accordance with standard procedures and recommendations . ApproprTwo single nucleotide polymorphisms (SNP), rs4341 and rs4343, previously found to be in complete linkage disequilibrium with the ACE I/D polymorphism were selected to design polymerase chain reaction-restriction fragment length polymorphisms (PCR-RFLP) . This me® cards and prepared for DNA analysis according to manufacturer's instructions [-1 buffer to water in an attempt to prevent any inhibition of the PCR by ethylenediamine tetraacetic acid. The PCR amplifications were implemented using one 2.0 mm punch from a Whatman Indicating FTA® card, 1 × GoTaq PCR buffer, 0.125 mM of each dNTP, 0.25 mM of each primer and 0.25 U GoTaq DNA polymerase  in a 10 μl reaction volume. The PCR conditions were as follows, 94°C for 3 min, 35 cycles of 94°C 30 sec, annealing for 30 sec and 72°C 30 sec, and a final extension at 72°C for 5 min in a PTC 200 thermocycler .Buccal cell samples were collected from participants using Whatman sterile foam tipped applicators and Whatman Indicating FTAructions . With thMethodology including annealing temperatures, genomic locations and primer sequences of ACE I/D and SNPs, as well as the restriction enzymes and their fragment sizes for PCR-RFLP tests and determination of ACE I/D genotype can be found elsewhere . DigestiDescriptive statistics were calculated for all variables. Sex differences were examined by independent t-test for continuous variables. Analysis of covariance (ANCOVA), controlling for age and sex, was used to examine differences in adiposity; and controlling for age, sex, and height to examine differences in BP variables across genotypes. These covariates were chosen since we included males and females in the analytic sample and the sample varied in age (3–12 years old). Height was included as a covariate in the analysis concerning BP since it is known to influence BP independent of age and sex. Post-hoc tests were performed to detect specific genotype group differences when a significant main effect for genotype was found from the one-way ANCOVA. The associations between adiposity and BP were examined using partial correlation coefficients, controlling for chronological age, sex, and height. The difference in the magnitude of the correlations between genotypes was tested using the Fischer z-transformation.To test the combined influence of ACE I/D genotype (II or D carrier) and body fatness  on BP, four groups were created: ACE II genotype and normal fat, ACE II and overfat, ACE D carriers and normalfat, and ACE D carriers and overfat. To test the combined influence, a one-way ANCOVA was conducted with BP measures  as the dependent variable testing for the main effect of combined genotype/body fatness group while controlling for the covariates age, sex, and height. Post-hoc tests were performed to detect specific genotype/fatness group differences when a significant main effect for combined genotype/body fatness was found from the one-way ANCOVA. A p-value of 0.05 was used for statistical significance. Statistical analyses were conducted using SPSS version 16.0.th percentile of the CDC growth charts. The mean BMI of boys approximated the 50th percentile, whereas the mean value for girls approximated the 75th percentile. There were significant sex differences for % body fat and BP, as girls had higher % body fat and DBP and MAP (p < 0.05). The frequency of ACE I/D genotypes did not differ by sex with the ID genotype being the most frequent (40.1%). The ACE I/D genotypes were not found to be in HW equilibrium (p < 0.022) for this sample of children.Table The physical characteristics of the subjects grouped by ACE I/D genotype are shown in Table Partial correlations between measures of adiposity and BP for the total sample and grouped by genotype are shown in Table In general, overfat children had higher BP than normal fat children (data not shown). Approximately 42% of subjects carrying the D allele were overfat compared to 33% in the normal fat group (p = 0.37). When the sample was analyzed by combined ACE I/D genotype and body fat groups, there was a significant main effect on SBP (p = 0.005)  , and a rAlthough the results of previous studies in youth are mixed, our results are generally consistent with the majority of these studies showing no significant association between ACE I/D genotype and BP in youth ,12,34,35Another purpose of this paper was to examine if the ACE I/D genotype modified the relationship between adiposity and BP. Our results showing a moderate correlation between adiposity and BP and differences between overfat and normal fat youth confirm previous work ,39. HoweThe role of ACE I/D genotype on adiposity and BP phenotypes of children are important to consider in the context of complex, multi-factorial phenotypes. First, these traits are not monogenic, and therefore other candidate genes influence these traits as well. Second, adverse exposure to other environmental factors may also be important to consider. However, we did show that BP was adversely influenced in the overfat D-carriers. Finally, it is possible that ACE I/D may influence these traits differently at various lifestages. Given the paucity of data in the area of genetics and pediatric health and the relative importance of understanding the role of the genome in human health and disease, additional study is warranted in this emerging field of study.The authors declare that they have no competing interests.JCE has made substantial contributions to all aspects of this paper including acquisition of funding; conception and design, acquisition of data, analysis and interpretation of data; and writing the manuscript. MAS has made contributions to the analysis and interpretation of data, and assisted in drafting the manuscript and revising it critically for important intellectual content. KG carried out the molecular genetic studies and drafted the methods section for the genotyping. MFR provided insight into the conception and design of the study, and provided critical feedback to the manuscript for important intellectual content. KAH was responsible for Acquisition of funding, coordinating data collection, and provided critical feedback to the manuscript for important intellectual content. All authors read and approved the final manuscript.