Sodium thiosulfate (STS) reduced calcium stone formation in both humans and genetic hypercalciuric stone forming (GHS) rats. We sought to measure urine chemistry changes resulting from STS administration in people.STS was given to healthy and hypercalciuric stone forming adults. Five normal non-stone forming adults (mean age 33 years), and 5 people with idiopathic hypercalciuria and calcium kidney stones (mean age 66 years) participated. Two baseline 24-hour urine collections were performed on days 2 and 3 of 3 days of self-selected diets. Subjects then drank STS 10 mmol twice a day for 7 days and did urine collections while repeating the self-selected diet. Results were compared by non-parametric Wilcoxon signed rank test. The primary outcome was the resulting change in urine chemistry.3 concentration after the STS period: it did not change. The net effect was an increase in supersaturation of uric acid, and no change in supersaturation of calcium oxalate or calcium phosphate.STS administration did not cause a significant change in urinary calcium excretion in either group. In both groups, 24 hour urinary ammonium (P = 0.005) and sulfate excretion (P = 0.007) increased, and urinary pH fell (P = 0.005); citrate excretion fell (P<0.05) in hypercalciuric participants but not in non-stone formers. Among stone formers with hypercalciuria, 3 of 5 patients had measurement of serum HCO3 did not change, urine tests suggested an acid load in both non-stone forming and hypercalciuric stone-forming participants. The long term safety of STS needs to be determined before the drug can be tested in humans for long-term prevention of stone recurrence.The basis for studies demonstrating that STS prevented stones in rats and people was not reflected by the changes in urine chemistry reported here. Although serum HCO Sodium thiosulfate (STS) has a number of uses in medicine, the most well known of which is its role as an antidote for cyanide poisoning. It acts via the conversion of cyanide to the more water-soluble thiocyanate which is readily excreted in urine Two studies, one in genetic hypercalciuric stone-forming (GHS) rats and one in humans, have shown a decrease in the incidence of kidney stones following STS administration Our study sought to investigate whether urine chemistries of humans on STS therapy would suggest efficacy in reducing urinary lithogenicity, and whether evidence for an associated acid load would be present.The study population consisted of two distinct arms. 2 (range 60.0- 91.0 ml/min/1.73 m2). Three patients had calcium stones composed of calcium oxalate and calcium phosphate, the latter component constituting 20, 40, or 50% of analyzed stones. The other 2 patients did not provide stones for analysis but had radio-opaque stones seen on plain abdominal radiography. Patients over the age of 80 were excluded from the study. Patients who had hypercalciuria in the past but did not have hypercalciuria on baseline collections done immediately prior to STS administration were not excluded. All patients were prescribed a diet demonstrated in a randomized controlled trial to reduce the incidence of recurrent stones in participants with hypercalciuria . The second group consisted of an equal number of participants with a mean age of 66 years. This group included 4 men and 1 woman all of whom had been seen and evaluated as outpatients in the Metabolic Stone Clinic and who were documented calcium stone formers with a history of hypercalciuria, specifically with 24 hour urine calcium measurements of greater than 300 mg/d on collections done within a year prior to the study, while following a self-selected diet. We identified 10 eligible patients seen in the Metabolic Stone Clinic in recent months and the first 5 who were contacted and agreed to participate were enrolled. The participants were recruited between January-April, 2011. The mean serum creatinine concentration was 1.1+0.1 mg/dl (range 0.9–1.2 mg/dl) and the mean estimated glomerular filtration rate was 74.2+12.4 ml/min/1.73 mThe patients were selected at random and solicited for participation. The study was approved by the institutional review board of the New York Harbor VA Healthcare System and registered with Clinicaltrials.gov as NCT01088555. The study was performed under US Food and Drug Administration Investigational New Drug number 106,424. All subjects provided written consent to participate in the study.Each participant undertook 2 baseline 24 hour urine collections over two consecutive days during which they were instructed to keep detailed food diaries while eating their self-selected diets. Oral doses of STS were then taken by the subjects for 7 days after the completion of the baseline urine collections. STS was dispensed by the research pharmacist to the patients and taken home for self-adminstration. The dose was 10 mmol (5 ml of a 2 M solution) twice a day, replicating the protocol described by Yatzidis. The taste and smell of STS precludes blinding the study. On days 5, 6 and 7, subjects replicated the diets they had documented during the pre-STS baseline urine collections. On days 6 and 7 of the medication administration period, participants undertook a second set of two 24 hour urine collections. Food diaries were not collected; they were simply used to allow subjects to replicate diets. Three of the 5 stone forming participants also had basic metabolic panels, including serum bicarbonate levels, checked prior to STS administration and on day 7 of taking STS; the other 2 did not have post-STS serum measurements because of logistical problems.During the urine collections, the urine was maintained at room temperature. An antimicrobial and a urine volume marker were added to each urine container and then a 50 ml aliquot of urine was obtained. The participants performed the urine collections at home, and then mailed their urine collections to Litholink Corp  for analysis In each 24 hour urine sample, we measured calcium, chloride, creatinine, magnesium, sodium, potassium, phosphate, ammonium, and uric acid concentrations by standard laboratory technique using a Beckman Synchron CX5 . pH was measured by glass electrode. Oxalate was measured by enzyme assay using oxalate oxidase . Citrate was measured by enzyme assay using citrate lyase . From these analyses supersaturation (SS) was calculated with respect to calcium oxalate, calcium phosphate, and uric acid using the iterative computer program EQUIL 2 Thiosulfate and sulfate were measured by ion chromatography using a Dionex ICS 2000 system . Samples were loaded into a 25-µl loop using an autosampler and injected onto an AG-11 guard column and AS-11 analytical column in series, with KOH as the mobile phase. Ion peaks were detected using a conductivity meter with the eluent background conductivity suppressed using an anion self-regenerating suppressor. We have found thiosulfate to be stable in urine under the collection conditions used in this study.P<0.05.Results of baseline and post-STS urine collections were compared by non- parametric Wilcoxon signed rank test and the effects of STS in controls and hypercalciuric stone forming participants were compared by Mann Whitney U test. Statistical data were generated and analyzed with a commercially available software package, Systat . Results were expressed as mean±SD and differences were considered statistically significant at All 10 patients enrolled completed the study without any lost to follow-up. The results of all participants were included. The results of STS administration in both the normal participants, the hypercalciuric participants and the 2 groups combined are presented in There was no evidence to suggest the development of metabolic acidosis based on the pre- and post-STS serum bicarbonate levels in the 3 stone forming patients from whom these data were collected . All 3 hUrine volume, creatinine, urea, and sodium excretion remained unchanged in both arms of the study pre- and post-STS, and in the combined data, confirming replication of diets over the course of the collections. Administration of the drug was well tolerated with no patients stopping therapy due to adverse events. All patients were questioned about their experience with the drug: 2 patients had transient episodes of loose or watery stools which resolved despite continued STS administration and 2 patients described foul smelling stool. No deviations from the protocol occurred.3]. Similar results occurred in the control non-stone-formers, with the exception that citrate did not fall. These effects on urine chemistry, with no net change in lithogenicity, would not explain remarkable reductions in stone activity reported in a non-controlled, non-randomized clinical study In the first study to examine a full set of urinary analytes after administration of oral STS, neither healthy controls, nor hypercalciuric calcium stone formers, experienced statistically significant increases in urine calcium excretion or calcium oxalate or calcium phosphate supersaturation. In stone-formers, there were decreases in urine pH and citrate, and increases in ammonium excretion without a change in serum [HCOOur urine results are also not consistent with the findings of Yatzidis et al 3] measurements while taking STS, all 3 had clear evidence of having received an acid load based on urine chemistry while maintaining normal serum [HCO3]. The lack of change in serum [HCO3] in the stone formers and the lack of change in urine citrate in the controls (in whom serum [HCO3] was not measured) suggest that ammoniagenesis was adequate to compensate for the presumed acid load. The difference in citrate excretion between controls and stone-formers despite similar increases in urine ammonium excretion, is interesting but not explained and suggests that perhaps stone-formers have greater renal citrate reabsorption in response to acid loads, contributing to relative hypocitraturia. The decrease in urine pH caused a significant increase in supersaturation of uric acid and could be associated with an increase in uric acid stone formation in some susceptible patients.The present study is consistent with thiosulfate presenting a net acid load, suggested by increases in urine ammonium excretion, and decreases in citrate excretion (in stone-formers) and urine pH. The etiology of this acid load is not clear. Stool losses of bicarbonate or potential base could be an explanation, though most patients had no diarrhea and those who did, experienced it only transiently. Previous reports of metabolic acidosis after parenteral administration of STS for cyanide poisoning and calcific uremic arteriolopathy exist a’s of that strong acid are 0.6 and 1.7 and the common knowledge that the sodium salt of a strong acid is not itself an acid. More likely, the acid load is due to the oxidation of thiosulfate to sulfate by the liver, producing protons Some recent literature has claimed that sodium thiosulfate is a strong acid The urine findings are similar to those noted in GHS rats given STS 3] did not change in the 3 in whom it was measured despite their significant changes in urine chemistry, we do not believe this small sample is an important limitation to knowing whether an acid load was presented. The urine findings clearly show that it occurred. We cannot speculate regarding the generalizability of these findings to a broader population of calcium stone formers with these or other urinary risk factors for stone formation.One limitation of the study is the small sample size of the 2 groups and that the groups are not matched for age. Since the primary outcome was not a comparison of the effects of the drug on the two populations we do not consider the age difference to be important. In fact, combining the results of the 2 groups demonstrated that the results were highly consistent between them with the only difference being a fall in urine citrate in hypercalciuric stone-formers but not in the non-stone-forming controls. Finding healthy people willing to take an investigational drug and do multiple urine collections among a hospital population was more difficult than finding stone formers interested in advancing the knowledge of their disorder and its potential treatment. Only after the results in the healthy group did we realize that it would be desirable to measure serum chemistry after drug administration and because of logistical issues could not get blood while participants were still taking the drug in 2 of the 5 stone formers. Since serum [HCO3] does not necessarily confirm the safety of long-term STS administration, as the finding is potentially related to two limitations of the study: its small sample size and relatively short duration of STS administration. STS therapy for the prevention of nephrolithiasis would likely need to continue for many years, and as such, studies of at least months to years would be necessary to sufficiently rule out development of sustained metabolic acidosis. A larger study with a longer duration more similar in scale and length to that of Yatzidis The absence of a significant change in serum [HCOThe appropriate therapy for prevention of recurrent calcium phosphate stones is unclear as no randomized controlled trials with this outcome have been completed


DNA repair gene X-ray repair cross complementing group 1 (XRCC1) plays an important role in the maintenance of the genomic integrity and protection of cells from DNA damage. Sequence variation in XRCC1 gene may alter head and neck cancer (HNC) susceptibility. However, these results are inconclusive. To derive a more precise estimation of the relationship between XRCC1 polymorphism and HNC risk, we undertook a meta-analysis involving 16,344 subjects.A search of the literature by PubMed, Embase, Web of Science and China National Knowledge Infrastructure was performed to identify studies based on the predetermined inclusion criteria. The odds ratio (OR) with 95% confidence interval (CI) was combined using a random-effects model or a fixed-effects model.Twenty-nine studies consisting of 6,719 cases and 9,627 controls were identified and analyzed. Overall, no evidence of significant association was observed between XRCC1 Arg194Trp, XRCC1 Arg280His, XRCC1 Arg399Gln genotypes and the risk of HNC in any genetic models. Subgroup analyses according to ethnicity, tumor site, publication year, genotyping method also detected no significant association in any subgroup, except that oral cancer was associated with Arg194Trp variant in recessive model. Furthermore, no significant effect of these polymorphisms interacted with smoking on HNC risk was detected but Arg194Trp homozygous variant.In conclusion, this meta-analysis suggests that the XRCC1 Arg194Trp, Arg280His and Arg399Gln polymorphism may not involve in HNC susceptibility. Further studies about gene-gene and gene-environment interactions in different populations are required. Head and neck cancer (HNC) is now the fifth most common type of cancer in the world , with apThe DNA repair system aims to maintain genomic integrity, and constantly challenge the environmental insults and replication errors. Therefore, the alteration of DNA repair genes could increase the risk of carcinoma in the head and neck . Three iMultiple studies have evaluated the association of HNC risk with polymorphism in the DNA repair genes XRCC1 Arg194Trp, XRCC1 Arg280His, and XRCC1 Arg399Gln. However, these results are inconsistent. While no association between XRCC1 polymorphisms and HNC risk was demonstrated in some studies ,12, but A systematic and electronic search of the PubMed, EMBASE, Web of Science, and China National Knowledge Infrastructure (CNKI) databases was performed to identify studies using combinations of the following search terms: “head and neck”, “oral”, “pharynx”, “larynx”, “nasopharynx”, “cancer”, “tumor”, “carcinoma”, “x-ray repair cross complementing group 1”, “XRCC1”, “Arg194Trp”, “Arg280His”, “Arg399Gln”, “polymorphism”, and “variation”. All of the studies were published from their earliest entry points to March 2013.All of the studies met the following inclusion criteria: (1) published in English; (2) examined case-control studies estimating the relationship between XRCC1 polymorphism and the risk of HNC; (3) described genotype frequencies; (4) genotype distribution in controls must be in Hardy-Weinberg equilibrium (HWE); and (5) when duplicated studies were published by the same author obtained from the same patient sample, only the most complete publication study was included in this meta-analysis. Unpublished reports and abstracts were not considered.The data were collected according to a standard protocol. The following information was extracted from each study: name of the first author, year of publication, country, genotyping methods, ethnicity and source of the cases and controls, characteristics of the sample population, and the genotype numbers from the cases and the controls.2 statistic [We first tested for deviations from the Hardy-Weinberg equilibrium (HWE) in the control groups using the goodness-of-fit test (Chi-square test or Fisher exact test). The odds ratio (OR) with a corresponding 95% confidence interval (CI) was used to examine the association between XRCC1 polymorphism and HNC risk. The current meta-analysis used the following statistical models, the allelic genetic model, the codominant genetic model (homozygote comparison), and the recessive genetic model. Heterogeneity among the studies was assessed using the chi-square-based Q statistic (P<0.1 for the Q test indicates significant heterogeneity) . We alsotatistic . Either tatistic ) or the tatistic ) was usetatistic ,22. P< 0As shown in In total, 29 reports, consisting of 6,719 cases and 9,627 controls, matching the inclusion criteria were included in the present meta-analysis. The characteristics are summarized in The distribution of XRCC1 Arg194Trp, XRCC1 Arg280His, and XRCC1 Arg399Gln polymorphism genotype frequencies between the HNC cases and the controls in the 29 studies are shown in The overall results of the meta-analysis for XRCC1 polymorphism and the risk of HNC are shown in A total of 22 studies, including 4,487 cases and 6,873 controls, examining the association between XRCC1 Arg194Trp polymorphism and HNC risk were reviewed. There was significant difference in the frequency of the XRCC1 Arg194Trp polymorphism between Caucasians and Asians . The pooled ORs for total population showed no evidence of a significant association between the variant genotypes of XRCC1 Arg194Trp and the risk of HNC in any genetic model. Significant heterogeneity was found in all genetic models. The forest plot is shown in Stratified analysis by ethnicity was performed in order to determine the source of heterogeneity among the studies. No significant association of HNC risk with XRCC1 Arg194Trp polymorphism was detected in Asians and Caucasians under any genetic model . SignifiOral cancer (OC) is the most common form of HNC and it is responsible for more than 90% of head and neck cancers . ConsequMany studies have demonstrated that the interaction between XRCC1 polymorphism and environmental toxins could influence the risk of HNC. Considering that smoking is a major aspect of environmental toxins, we performed a subgroup analysis of six studies to investigate the influence that the interaction of tobacco smoke with XRCC1 polymorphism has on HNC risk. There was a significant association between the joint effect of smoking with XRCC1 Arg194Trp polymorphism and the risk of HNC under homozygote comparison . No signFlores-Obando et al. conducted a similar meta-analysis that included studies published before 2010. Considering the inconsistent results between the two studies, we decided to perform a stratified analysis by including studies published after 2010. The result showed no significant association was detected between Arg194Trp polymorphism and HNC risk in any genetic model . BetweenThe different genotyping methods used in the literature could cause the different genotyping results. Therefore, we performed a subgroup analysis by genotyping methods to investigate the relationship between Arg194Trp and HNC susceptibility. Neither the PCR-RFLP subgroup nor the TaqMan subgroup detected any significant association in the analyses for all genetic models . MoreoveEleven studies, including 2,972 cases and 3,714 controls, examining the relationship between XRCC1 Arg280His polymorphism and HNC risk were reviewed. There was significant difference in the frequency of the XRCC1 Arg280His polymorphism between Caucasians and Asians . There was no significant association of HNC risk with variant genotypes of XRCC1 Arg280His in any genetic model. Significant between-study heterogeneity was absence in all genetic models. The forest plot is shown in We conducted subgroup analyses by ethnicity, tumor site, publication year, and genotyping method to estimate the relationship between XRCC1 Arg280His variant genotypes and the risk of NHC. However, no significant association was observed in any subgroup under different genetic models, and there was no significant heterogeneity among the studies in any stratified analysis . Only onThere were 27 studies, including 6,466 cases and 9,379 controls, that examined the association between HNC susceptibility and XRCC1 Arg399Gln polymorphism. There was significant difference in the frequency of the XRCC1 Arg399Gln polymorphism between Caucasians and Asians . Overall, the association between variant genotypes of XRCC1 Arg399Gln polymorphism and HNC susceptibility was not significant under the allelic genetic model, homozygote comparison, and the recessive model. Between-study heterogeneity was detected in all genetic models. The forest plot is shown in In the subgroup analysis by ethnicity, no significant association between XRCC1 Arg399Gln polymorphism and HNC risk was found in Asians  and CaucFour studies, including 663 cases and 1,162 controls, were performed on OC population, and there was no significant association between XRCC1 Arg399Gln polymorphism and HNC susceptibility . BetweenIn the stratified analysis by smoking in the allelic genetic model and homozygote comparison, four studies were included and no significant association was found . Seven sIn stratified analysis by publication year of literature published from 2010–2012, significant heterogeneity was detected in all genetic models. Moreover, we found no association between Arg399Gln and HNC risk under any genetic model .In subgroup analysis of genotyping method, PCR-RFLP, TaqMan, and sequence analysis were used in the literature for genotyping of XRCC1 Arg399Gln polymorphism. The results showed no significant association between Arg399Gln and HNC risk in any stratified analysis under different genetic models .Both Begg’s test and Egger’s test were performed to assess the publication bias of the literature. Visual analysis of the funnel plots did not present any evidence of obvious asymmetry for any genetic model in the overall meta-analyses of XRCC1 Arg194Trp, Arg280His, and Arg399Gln . HoweverDNA repair mechanisms play a critical role in the protection of cells from DNA damage and in the maintenance of genomic integrity. The protein encoded by the XRCC1 gene is a scaffolding protein that associates with DNA ligase I, DNA ligase III, polynucleotide kinase (PNK), DNA polymerase β, and poly polymerase, which are parts of the DNA repair system. The interaction of XRCC1 with DNA ligase III could increase the endocellular stability of ligase. The joint effects of XRCC1 and PNK stimulate the 5’-kinase and 3’-phospatase activities. All of these conditions promote the repair of DNA. Therefore, sequence variation in the XRCC1 gene is suggested to alter cancer’s susceptibility. The most common variant genotypes of XRCC1, including the Arg194Trp, Arg399Gln, and Arg280His genes, are described and a number of studies have investigated the genetic effect of the XRCC1 Arg194Trp, Arg280His, and Arg399Gln polymorphisms on HNC susceptibility with inconsistent results. This diversity motivates the current updated meta-analysis that may help us to explore a more robust estimate of the effect of XRCC1 polymorphism on the risk of HNC. In the present meta-analysis of 6,719 cases and 9,627 controls, no evidence of a significant association between HNC susceptibility and any type of XRCC1 variant genotype was detected.A previous meta-analysis, conducted in 2010 by Flores-Obando et al., evaluated the relationship between XRCC1 polymorphisms and the risk of HNC based on 15 publications including 2,330 cases and 3,834 controls for Arg194Trp, four publications including 879 cases and 926 controls for Arg280His, and 15 studies including 3,582 cases and 5,347 controls for Arg399Gln polymorphism. We updated this meta-analysis by adding the sample sizes. A total of 22 studies, including 4,487 cases and 6,873 controls, evaluated the association between the XRCC1 Arg194Trp gene and HNC risk; 11 studies, including 2,972 cases and 3,714 controls, evaluated the association between Arg280His and HNC risk; and 27 studies, including 6,466 cases and 9,379 controls, evaluated the association between Arg399Gln polymorphism and HNC risk. There are some discrepancies between the Flores-Obando et al. meta-analysis and ours. A marginal association between XRCC1 Arg399Gln polymorphism and HNC risk was detected under the recessive genetic model on Caucasians in the meta-analysis conducted by Flores-Obando et al., but it was not found in ours. These diverse results may, generally, be due to the differences in the studies included in the meta-analysis. Nasopharyngeal carcinoma (NPC) is one type of HNC, which has a striking geographic and ethnic distribution, with particularly high rates observed among Asians. The literature on NPC was not included in the Flores-Obando et al. study, but it was included in ours. In the above-mentioned stratified analysis result, seven articles were shared the Flores-Obando et al. study and our study, and our study included an additional seven articles, including newly published literature and NPC literature. The results of these seven studies account for 54.94% weight , which cThe relationship between HNC susceptibility and variant genotypes of XRCC1 might be affected by the tumor sites. Accordingly, we also performed stratified analysis in the oral cancer group. The result of this subgroup analysis showed no evidence for a significant association between XRCC1 polymorphism and the risk of OC in any genetic model, except for the Arg194Trp variant in the recessive model. These results generally agreed with the meta-analysis conducted by Zhou et al. , but theRecent studies have reported on the associated risk of XRCC1 polymorphism cross lifestyle factors in the progression of head and neck cancer. Cigarette smoking is a major subject of investigation in various cancers. In 1997, the World Health Organization reported that there were 1.1 billion smokers worldwide and smoking-related cancers accounted for 22% of all cancers. Hence, subgroup analysis to estimate the interaction between the genotypes of XRCC1 Arg194Trp and Arg399Gln and smoking on HNC risk was performed. The results showed that the interaction of smoking and Arg399Gln variant genotypes displayed no statistical significance in all three genetic models. A significant association between the joint effect of smoking and XRCC1 Arg194Trp polymorphism and HNC susceptibility was detected under a homozygote comparison , but no statistical significance was revealed in the other genetic models. In the forest plot of HNC risk associated with the interaction between smoking and Arg194Trp, the result of a study by Cao et al. accounted for 71.15% weight , which mBetween-study heterogeneity is a well-known problem that is unavoidable. In our meta-analysis, heterogeneity was detected in the Arg194Trp and Arg399Gln polymorphism total population groups . The souPublication bias is a well-known problem that was not found by funnel plot for the overall meta-analyses of the XRCC1 polymorphisms . HoweverSome limitations should be considered in our meta-analysis. First, some of the included studies in our meta-analysis contained a smaller sample size, which might result in a lack of ability to detect the possible risk for XRCC1 polymorphism. Second, due to the limited number of studies, subgroup analysis was not performed in Africans . Third, Despite of the limitations mentioned above, the results of the current meta-analysis suggest that XRCC1 Arg194Trp, Arg280His, and Arg399Gln polymorphism is not involved in HNC susceptibility. In addition, further studies evaluating the effect of gene-gene and gene-environment interactions on these gene polymorphisms with HNC susceptibility are required, especially in an African population.Figure S1Forest plot of HNC risk associated with XRCC1 Arg194Trp gene polymorphism under all genetic models on Asians.(DOC)Click here for additional data file.Figure S2Forest plot of HNC risk associated with XRCC1 Arg194Trp gene polymorphism under all genetic models on Caucasians.(DOC)Click here for additional data file.Figure S3Forest plot of HNC risk associated with XRCC1 Arg194Trp gene polymorphism under all genetic models in oral cancer population.(DOC)Click here for additional data file.Figure S4Forest plot of HNC risk associated with XRCC1 Arg194Trp gene polymorphism under all genetic models in the stratified analysis by studies published after 2010.(DOC)Click here for additional data file.Figure S5Forest plot of HNC risk associated with XRCC1 Arg194Trp gene polymorphism under all genetic models in using PCR-RFLP analysis.(DOC)Click here for additional data file.Figure S6Forest plot of HNC risk associated with XRCC1 Arg194Trp gene polymorphism under all genetic models in using TaqMan analysis.(DOC)Click here for additional data file.Figure S7Forest plot of HNC risk associated with XRCC1 Arg280His gene polymorphism under all genetic models on Asians.(DOC)Click here for additional data file.Figure S8Forest plot of HNC risk associated with XRCC1 Arg280His gene polymorphism under all genetic models on Caucasians.(DOC)Click here for additional data file.Figure S9Forest plot of HNC risk associated with XRCC1 Arg280His gene polymorphism under all genetic models in oral cancer population.(DOC)Click here for additional data file.Figure S10Forest plot of HNC risk associated with XRCC1 Arg280His gene polymorphism under all genetic models in the stratified analysis by studies published after 2010.(DOC)Click here for additional data file.Figure S11Forest plot of HNC risk associated with XRCC1 Arg280His gene polymorphism under all genetic models in using PCR-RFLP analysis.(DOC)Click here for additional data file.Figure S12Forest plot of HNC risk associated with XRCC1 Arg280His gene polymorphism under all genetic models in using TaqMan analysis.(DOC)Click here for additional data file.Figure S13Forest plot of HNC risk associated with XRCC1 Arg399Gln gene polymorphism under all genetic models on Asians.(DOC)Click here for additional data file.Figure S14Forest plot of HNC risk associated with XRCC1 Arg399Gln gene polymorphism under all genetic models on Caucasians.(DOC)Click here for additional data file.Figure S15Forest plot of HNC risk associated with XRCC1 Arg399Gln gene polymorphism under all genetic models in oral cancer population.(DOC)Click here for additional data file.Figure S16Forest plot of HNC risk associated with interaction between XRCC1 Arg399Gln polymorphism and smoking under all genetic models.(DOC)Click here for additional data file.Figure S17Forest plot of HNC risk associated with XRCC1 Arg399Gln gene polymorphism under all genetic models in the stratified analysis by studies published after 2010.(DOC)Click here for additional data file.Figure S18Forest plot of HNC risk associated with XRCC1 Arg399Gln gene polymorphism under all genetic models in using PCR-RFLP analysis.(DOC)Click here for additional data file.Figure S19Forest plot of HNC risk associated with XRCC1 Arg399Gln gene polymorphism under all genetic models in using TaqMan analysis.(DOC)Click here for additional data file.Figure S20Forest plot of HNC risk associated with XRCC1 Arg399Gln gene polymorphism under all genetic models in using sequence analysis.(DOC)Click here for additional data file.Supplement S1PRISMA Flowchart.(DOC)Click here for additional data file.Supplement S2PRISMA Checklist.(DOC)Click here for additional data file.


P = 0.2367). Our findings therefore suggest no individual or interactive association between four prominent Hcy pathway markers and EH.Hyperhomocysteinemia (hHcy) has been associated with an increased risk of cardiovascular disease and stroke. Essential hypertension (EH), a polygenic condition, has also been associated with increased risk of cardiovascular related disorders. To investigate the role of the homocysteine (Hcy) metabolism pathway in hypertension we conducted a case-control association study of Hcy pathway gene variants in a cohort of Caucasian hypertensives and age- and sex-matched normotensives. We genotyped two polymorphisms in the methylenetetrahydrofolate reductase gene (MTHFR C677T and MTHFR A1298C), one polymorphism in the methionine synthase reductase gene (MTRR A66G), and one polymorphism in the methylenetetrahydrofolate dehydrogenase 1 gene (MTHFD1 G1958A) and assessed their association with hypertension using chi-square analysis. We also performed a multifactor dimensionality reduction (MDR) analysis to investigate any potential epistatic interactions among the four polymorphisms and EH. None of the four polymorphisms was significantly associated with EH and although we found a moderate synergistic interaction between MTHFR A1298C and MTRR A66G, the association of the interaction model with EH was not statistically significant ( Hypertension is defined as a sustained systolic blood pressure of greater than 140 mmHg, a diastolic blood pressure of greater than 90 mmHg, or both . In addiThere are a number of environmental and clinical risk factors associated with EH including, but not limited to, dietary intake of sodium, alcohol intake, lack of exercise, poor diet, obesity, insulin resistant diabetes, and hyperlipidemia. Although these factors explain a substantial proportion of hypertension susceptibility, it is estimated that up to 60% of the variation in hypertension risk is due to an individual's genetic makeup . Thus, mThe homocysteine (Hcy) pathway has emerged as a strong candidate for EH and many studies have investigated genetic variation underlying hyperhomocysteinemia (hHcy). However, results have so far been inconclusive, with some studies reporting a significant association –10 whileOne of the most studied genetic variants contributing to hHcy is the C to T single nucleotide polymorphism (SNP) at codon 677 of the MTHFR gene. The C to T substitution causes alanine to be substituted by valine. The TT variant codes for a thermolabile enzyme which has a 50% reduced activity compared to the CC variant . AnotherAlthough the MTRR enzyme does not directly participate in the conversion of Hcy to methionine, the fact that it keeps the MTR enzyme active makes it a key enzyme in Hcy metabolism. A common SNP in MTRR is the A to G substitution at codon 66. This substitution causes isoleucine to be substituted by methionine in the enzyme. It has been reported that the mutant enzyme exhibits a four-fold lower activity in reactivating MTR than the wild type enzyme . This poThis study investigated whether there is an association between EH and the MTHFR C677T, MTHFR A1298C, MTRR A66G, and MTHFD1 G1958A variants in an Australian case-control cohort. An interaction analysis using the multifactor dimensionality reduction (MDR) method was also performed to investigate whether specific combinations of genotypes across all four loci contribute to disease status.MDR analysis is a data mining method used to detect and classify combinations of independent variables such as genotypes or environmental factors that may interact to cause disease. MDR classifies the genotype combinations of two loci (multilocus genotype) into either belonging to a low-risk group or a high-risk group. For example, all possible genotypes at locus 1  are paired with each other possible genotype at locus 2 , giving nine possible multilocus genotypes . Each multilocus genotype is then evaluated for the number of cases versus controls, and assigned to be high-risk if the number of cases exceeds the number of controls, corresponding to a ratio >1 for matched populations , 22. If The study protocol was approved by the Griffith University's Ethics Committee. The study population was composed of 409 hypertensives and 409 age- (±5 years), sex-, and ethnicity-matched normotensive controls, who resided in the South East Queensland region of Australia. All participants were of Caucasian origin. Cases were defined as individuals who were clinically diagnosed as suffering from hypertension and who were taking antihypertensive drugs. Controls were defined as participants who were not taking antihypertensive drugs, and whose blood pressure was less than 140/90 mmHg. Individuals suffering from renal disorders , primary aldosteronism, Cushing syndrome, and hypothyroidism were excluded from the study. None of the participants included in the study reported any previous cardiovascular events such as heart attacks or stroke. 53.3% of the population were female and 46.7% were male. The average age of the case group was 63.1 ± 10.9 years and the average age of the control group was 61.0 ± 10.5 years. Peripheral blood samples as well as questionnaires detailing medical history, including blood pressure and prescribed medications, were obtained from all participants. All participants signed informed consent agreements prior to collection of blood and clinical information.2, GoTaq polymerase were from Promega Corp., Madison, WI, USA; dNTPs, restriction enzymes, and enzyme buffers were from New England Biolabs, Ipswich, MA, USA; SYTO9 dye was from Invitrogen, Carlsbad, CA, USA. Protocol and assays for each polymorphism are described in detail below.DNA was extracted from blood samples using a modified version of the salting-out method . Two pol2, 0.2 mM dNTPs, 0.2 uM forward primer, 0.2 uM reverse primer, 1U GoTaq, and 40 ng of DNA. The primer sequences were designed by Frosst [HinfI and 1X NEB Buffer 2 at 37°C for 12 hrs, followed by an 80°C enzyme deactivation step of 20 mins. Restriction digest products were electrophoresed on a 15 cm 3.5% agarose gel for 120 min at 80 V, which was then poststained in a 0.01% solution of ethidium bromide in 1X TAE buffer for 40 min and visualised under ultraviolet light. HinfI digestion of fragments containing the T allele produced two fragments of 175 bp and 23 bp while fragments containing the C allele remained undigested by HinfI.The MTHFR C677T polymorphism was genotyped by polymerase chain reaction (PCR) followed by restriction fragment length polymorphism (RFLP) analysis. The PCR protocol was as follows: 1X PCR buffer, 1.75 mM MgCly Frosst  and werey Frosst . The PCR2, 0.2 mM dNTPs, 0.3uM forward primer, 0.3uM reverse primer, 1.6uM SYTO9, 1U GoTaq, and 40 ng of DNA. The primer sequences were obtained from a previous study [The MTHFR A1298C polymorphism was genotyped by PCR followed by high resolution melt (HRM) analysis. The PCR protocol was as follows: 1X PCR buffer, 1.5 mM MgClus study  and wereus study . The PCR2, 0.2 mM dNTPs, 0.3 uM forward primer, 0.3 uM reverse primer, 1.6 uM SYTO9, 1U GoTaq. The primer sequences were obtained from a previous study [The MTRR A66G polymorphism was genotyped by PCR followed by HRM analysis. The PCR protocol was as follows: 1X PCR buffer, 1.5 mM MgClus study  and wereus study . The PCR2, 0.2 mM dNTPs, 0.3 uM forward primer, 0.3 uM reverse primer, 1.6 uM SYTO9, 1 U GoTaq. The primer sequences were obtained from a previous study [The MTHFD1 G1958A polymorphism was genotyped by PCR followed by HRM analysis. The PCR protocol was as follows: 1X PCR buffer, 1.5 mM MgClus study  and wereus study . The PCRPower analysis for this study was performed using the Power for Genetic Analyses software . GenotypGiven the possibility that each variant may only contribute a small independent effect which may not be detectable as statistically significant in our case control cohort, we also performed interaction analysis using the MDR 2.0 software version beta 8.4. The MDR program was designed to test for interactive genetic effects on a trait even if the independent effects are nonsignificant .In the MDR software, main effect (one-locus) models, two-locus models, or N-locus models are generated, and each model is assessed for prediction accuracy by dividing the dataset into multiple sets, with one set excluded from model-training and then used to test the model. The process of division, model-training, and model-testing is repeated multiple times to cross-validate each model. Testing accuracy (TA) and cross-validation consistency (CVC) are then used to evaluate the overall best model. Permutation testing can then be performed on the dataset using an additional module called MDRpt, which evaluates the significance of the model TA .r2 > 0.85) and all were used in the MDR analysis. Missing genotypes were then imputed by mode substitution. Software default settings were used except that the cross-validation was repeated 100 times, and paired analysis was selected. The model with the highest TA and CVC was determined to be the best model and significance P values were then generated using 10,000 permutations in the MDR permutation testing module (MDRpt) version 1.0 beta 2.Before performing the MDR analysis, all markers were examined for correlation using PLINK's pairwise LD function , to idenP > 0.05).This study has more than 90% power to detect a relative risk of at least 1.5 for all markers. Genotype and allele frequencies for all four markers are shown in χ2 = 0.03, P = 0.99) or the A1298C marker . There was also no statistically significant difference between the allele frequencies of cases and controls for either the C677T  or the A1298C  polymorphisms. For the C677T marker, there was no observed trend in either the genotype or allele frequencies, with the TT genotype frequency at 8.7% for cases and 8.9% for controls, and the T allele frequency at 31.8% for cases and 32.2% for controls. For the A1298C marker, there was an increased AA genotype frequency in cases (44.8%) compared to controls (42.0%), though this trend was less apparent in A allele frequency in cases (65.4%) compared to controls (64.4%). The observed minor allele frequencies in the control group for both the C677T marker  and the A1298C marker  conformed well with expected control frequencies for each marker  as determined in the Hap-Map CEU population (Utah residents of Northern European ancestry).For MTHFR, there was no statistically significant difference between the genotype frequencies of cases and controls for either the C677T marker , or the allelic frequencies of cases and controls . The GG genotype frequency was 18.1% for cases and 20.7% for controls, while allele frequencies showed a trend of decreased G allele frequency in cases (44.7%) compared to controls (47.1%). Although the genotype frequencies of our control group seemed markedly different to the Hap-Map CEU frequencies with 52.8% of heterozygotes in our control population compared to only 34.0% in the Hap-Map CEU population, the allelic frequencies of our control group  and the Hap-Map CEU population  were similar.Similarly, for the MTRR A66G polymorphism, there was no statistically significant difference between either the genotype frequencies of cases and controls  or the allelic frequencies of cases and controls . The GG genotype frequency was 32.7% for cases and 28.9% for controls, while the G allele frequency was 55.8% for cases and 54.3% for controls. The observed allele frequencies for our control group  was similar to expected allele frequencies as determined by the Hap-Map CEU population . Case and control genotype frequencies were also analysed by gender .For the MDR analysis, the best MDR models for the one SNP (main effect), two SNP, and three SNP combinations are shown in We investigated the homocysteine pathway variants MTHFR C677T, MTHFR A1298C, MTRR A66G, and MTHFD1 G1598A in an Australian Caucasian population for association with EH. There was no statistical difference between our case and control groups for either genotype or allele frequencies for any of the markers studied, indicating no detected association between these four markers and EH in our case-control population. However, given the sample size limitation, we could not rule out the possibility that these variants contributed a modest effect on EH in this cohort (OR < 1.5) that was not detectable as statistically significant in this study, therefore, we conducted the interaction analysis using an MDR approach. We found that the best model indicated an interaction between the two SNPs MTHFR A1298C and MTRR A66G, which was found to be nonsignificant by permutation testing. This may reflect the fact that the mechanism by which hHcy can cause hypertension is not well understood. However, a recent study in human umbilical artery smooth muscle cells reported an increase in the proliferation of vascular smooth muscle cells through the Hcy-mediated differential regulation of cyclin A and D1, which led to an increase in intima media thickness . AnotherMTHFR has been among the most studied genes in relation to Hcy and folate metabolism, with regard to a variety of diseases ranging from neural tube defects to CVD and EH. Previous studies have shown that the MTHFR variants C677T and A1298C have been associated with both higher levels of Hcy  and EH rAnother meta-analysis of Hcy metabolizing enzymes and risk of coronary heart disease consisting of 23 studies reported an association of the C allele of the MTHFR A1298C with myocardial infarction with an OR of 1.37 (95% CI 1.03–1.84) . HoweverMTRR and MTHFD1 have both been shown to carry variants which decrease enzymatic activity and disrupt either MTR reactivation (for MTRR) or purine synthesis (for MTHFD1) though MTHFD1 has not been previously studied in relation to EH. The MTRR A66G polymorphism has been associated with increased Hcy levels . HoweverCurrent data from this and other studies suggest that genes within the Hcy pathway are not significantly associated with an increase in EH risk, including the well-studied marker MTHFR C677T. Additionally, given that each gene may confer a modest effect to EH risk, a polygenic profile analysis of genes in the Hcy pathway may be warranted. Additionally, future studies should measure plasma Hcy levels to determine whether a combination of these markers influences Hcy levels overall. Our study could not verify whether Hcy levels are significantly different between our cases and controls, and whether individual markers or combinations of markers influence EH risk through elevating Hcy levels. Further, the effects of diet on Hcy levels and EH risk should be controlled for in any future analysis as protective diet such as high folate intake may abrogate an increased genetic risk to EH due to genetic variations in the Hcy pathway.


HLA-DRB1 allele polymorphisms have been reported to be associated with hepatocellular carcinoma susceptibility, but the results of these previous studies have been inconsistent. The purpose of the present study was to explore whether specific HLA-DRB1 alleles  confer susceptibility to hepatocellular carcinoma.Case-control studies on HLA-DRB1 alleles association with HCC were searched up to January 2010 through a systematic review of the literature. The odds ratios (ORs) of HLA-DRB1 allele distributions in patients with hepatocellular carcinoma were analyzed against healthy controls. The meta-analysis software REVMAN 5.0 was applied for investigating heterogeneity among individual studies and for summarizing all the studies. Meta-analysis was performed using fixed-effect or random-effect methods, depending on absence or presence of significant heterogeneity.OR = 2.88, 95%CI: 1.77-4.69, P <0.001, respectively).Eight case-control studies were included in the final analysis. Among the 3 HLA-DRB1 alleles studied, DRB1*07 and DRB1*12 were significantly associated with the risk of HCC in the whole populations . No significant association was established for DRB1*15 allele with HCC in the whole populations. Subgroup analysis by ethnicity showed that DRB1*07, DRB1*12 and DRB1*15 alleles significantly increased the risk of hepatocellular carcinoma in Asians (OR = 2.10, 95% CI: 1.06-4.14, P = 0.03; OR = 1.73, 95% CI: 1.17-2.57, P = 0.006 and These results support the hypothesis that specific HLA-DRB1 alleles might influence the susceptibility of hepatocellular carcinoma. Large, multi-ethnic confirmatory and well designed studies are needed to determine the host genetic determinants of hepatocellular carcinoma. Hepatocellular carcinoma (HCC) is linked to the interaction between genetic, immunologic, environmental, dietary, and life style factors. Its incidence and distribution vary widely among ethnic groups, sex, and geographic regions. Hepatocellular carcinoma is the third most common cause of cancer-related deaths worldwide with about 600,000 patients dying from the disease annually . Asian cThe Major Histocompatibility Complex (MHC) plays a key role in anti-virus and tumor defense. Human leukocyte antigens (HLA) function in the regulation of immune response to foreign antigens and discrimination of self from non-self antigens. They are encoded by a series of closely linked genetic loci found on chromosome 6 ,5. HLA pElectronic databases  were used to search for all genetic association studies evaluating the HLA-DRB1 polymorphism and HCC in humans in all languages up to January 2010. The search strategy was based on combinations of the terms: HLA-DRB1 AND (Hepatocellular carcinoma or HCC) AND . We also performed a full manual search from the bibliographies of selected papers. We also contacted the authors of studies containing relevant information, who did not report the results necessary for this analysis. Unpublished data were also accepted if an abstract was available and further information was obtained from the author.In the meta-analysis, the following inclusive selection criteria were set and reviewed by two independent investigators: (1) each trial is an independent case-control study; (2) the purpose of all studies and statistical methods is similar; (3) it supplied enough information to calculate the odds ratio (OR);(4) HLA-DRB1 alleles were molecularly typed (high or low resolution level); (5) inclusion of patients according to the diagnosis standard of HCC defined in 2002, based on at least one of the following criteria: classical histological characteristics or serum a-fetoprotein (AFP) level higher than 400 ng/ml together with radiological findings (ultrasound and/or CT) consistent with HCC . A singlAlthough assessment of study quality is considered important for systematic reviews and meta-analyses, scoring methods have been considered problematic  and may The frequency of HLA-DRB1 alleles varies according to ethnic and racial background, with some alleles being extremely rare. Therefore, articles were not required to identify all alleles for inclusion.The studies were independently evaluated by two researchers. Discrepancies in the evaluations of some studies were resolved by discussion between the reviewers. The following data were collected from each study: authors, publication year, journal, publication type and language, HLA genotyping method, allele genotyped, allele frequencies, numbers of cases and controls, definitions criteria used for HCC, HCC sample description, controls sample description. Allelic frequency was calculated as the number of cases or controls harboring at least one allele type (HLA-DRB1) divided by the total number of chromosomes included in each of the corresponding groups.2I) tests. 2 Ivalues of 25%, 50% and 75% were assigned as low, moderate, and high estimates, respectively. If the results of the Q test had no significant heterogeneity, the Mantel-Haenszel fixed effect model (Peto method) was used for the combination of data; If the results of the Q test had significant heterogeneity, the Dersimonian-Laird random effects model (DL method) were used for the combination of data [absence or presence of significant heterogeneity. As a measure of association between HCC and HLA-DRB1 alleles, we combined odds ratios (ORs) with 95% confidence intervals (95%CIs) stratified by gene subtype of patients and controls in a study. Funnel plots and the Egger's regression asymmetry test were used to evaluate publication bias [P values presented are two-tailed. To reduce heterogeneity and to evaluate whether there was a different genotype effect in predefined subgroups of studies, we performed subgroup analysis according to the ethnicity. We performed a sensitivity analysis to assess the stability of the results by sequential omission of individual studies. The analyses were performed using Revman 5.0 provided by the Cochrane Collaboration.The literature review conformed to PRISMA statement standards, and our research fit the minimum set of items for reporting in systematic reviews and meta-analyses . Five studies used low resolution molecular typing for HLA, while three used high resolution molecular typing for HLA. Low resolution molecular typing methods for HLA could not identify the specific alleles. Accurate methods for HLA class II typing should involve the combination of PCR-SSOP, PCR-SSP, and PCR-SSCP .A total of 13 HLA-DRB1 alleles were studied in the 8 case-control studies, but only 3 alleles  was extracted from the studies to investigate their association with HCC, which were reported at least six of the eight case-control studies. Statistics calculated for each study are shown in the forest plot Figures .2 = 10.79, 2 I= 54%, P = 0.06 and >0.05). Under the fixed effect model, the combined OR for the association of HLA-DRB1*07 allele with the risk for HCC in the whole populations was determined to be 1.65 , and under the random effects model was 1.77 . In sensitivity analysis, the exclusion of individual studies did not change this significant result, except for exclusion of the study by Pan et al and El-Chennawi et al, which produced a non-significant association. Subgroup analysis by ethnicity showed that HLA-DRB1*07 allele significantly increased the risk of hepatocellular carcinoma in Asians under the fixed effect model .In the meta-analysis, overall the frequencies of HLA-DRB1*07 allele was 20.1% (59 of 281) in HCC and 15.2% (71 of 466) in controls. The heterogeneity test indicates that the variation of trial-specific ORs was not statistically significant , so the fixed-effect method was used to combine the results. The combined OR for the association of HLA-DRB1*12 allele with the risk for HCC in the whole populations was determined to be 1.59 (95% CI: 1.09-2.32), and was statistically significant (P = 0.02 and <0.05). In sensitivity analysis, the exclusion of individual studies did not change this significant result, except for exclusion of the study by Donaldson, et al and Sun, et al, which produced a non-significant association. Subgroup analysis by ethnicity showed that HLA-DRB1*12 allele significantly increased the risk of hepatocellular carcinoma in Asians .Overall the frequencies of HLA-DRB1*12 allele was 26.3% (74 of 281) in HCC and 16.1% (83 of 516) in controls. The heterogeneity test indicates that the variation of trial-specific ORs was not statistically significant .Meta-analysis for HLA-DRB1*15 allele was carried out, but it did not show any statistical effect in the whole populations. Subgroup analysis by ethnicity showed that HLA-DRB1*15 allele significantly increased the risk of hepatocellular carcinoma in Asians under the fixed effect model .To our knowledge, this is the first published meta-analysis investigating this association. Our meta-analysis of eight studies revealed that DRB1*07 and DRB1*12 were significantly associated with the risk of HCC in the whole populations . Subgroup analysis by ethnicity showed that DRB1*07 and DRB1*12 alleles also significantly increased the risk of hepatocellular carcinoma in Asians . No significant association was established for DRB1*15 allele with HCC in the whole populations. Subgroup analysis by ethnicity showed that DRB1*15 alleles significantly increased the risk of hepatocellular carcinoma in Asians under the fixed effect model  reported globally each year, and Hepatitis B Virus (HBV) is the major etiology of HCC in these areas. Although HBV infection plays an important role in HCC, HBV infection alone is not sufficient for progression to HCC. Several lines of evidence suggest that cellular immune surveillance is important in the control of HBV infection and the development of HCC. In 2007, Yang and his colleagues found that HLA-DRB1*07 were markedly higher in the HBV-infected group among people in northwestern China  . In 2006Moreover, HLA-DRB1 alleles polymorphism have been reported to be associated with other cancers and autoimmune diseases, including cervical squamous cell carcinoma, rheumatoid arthritis, systemic lupus erythematosus, autoimmune hepatitis, inflammatory bowel disease, multiple sclerosis and type 1 diabetes, and meta-analyses have been done for these diseases -31.Additionally, because the information used in our research was based on data from observational studies, some limitations should be discussed in this meta-analysis and the results should be considered with caution. A primary cause for the difference in results by different authors may be related to the great variability of the frequency of HLA alleles in different populations. It is quite possible that one ethnic group may have some specific alleles in development/protection of HCC compared to other ethnic groups. Many studies were conducted on relatively smaller samples. Insufficient number of individuals might decrease the power to detect a difference in the distribution of DRB1*07, DRB1*12 and DRB1*15 alleles between HCC patients and controls, though a true difference exist. Lack of an association may not mean that associations do not exist. Many studies did not control for the matching variables in the analysis, and the possible confounders could be among the potential causes of variation in the studies' estimates. Different types of control groups could also be among the potential causes of variation in the studies' estimates. Effects of interactions in other environmental/behavioral and/or viral factors may be inevitable. A complex interplay between various genes is likely to modulate the development of HCC rather than a single allele. HLA genotyping techniques must be taken into consideration because these methodologies have different sensitivities. HLA genotyping were mainly tested for by PCR with probe hybridization, but targets were not always verified by sequencing. This could hinder an effective comparison between the studies and influence the combined results. The shape of the funnel plot seemed to be asymmetrical, suggesting that publication bias might affect the findings of our meta-analysis. Furthermore, although we tried to maximize our efforts to identify all relevant published studies in peer-reviewed journals, it is possible that some escaped our attention.In spite of these, our meta-analysis also had some advantages. First, substantial number of cases and controls were pooled from different studies, which significantly increased statistical power of the analysis. Second, the quality of case-control studies included in the current meta-analysis was satisfactory based on our selection criteria. Third, the patients were phenotypic homogeneous subjects, diagnosed either by gold standard or by biochemical and imageology combined methods, which could reduce the heterogeneity in some extent.Our meta-analysis suggests that HLA-DRB1*07 and DRB1*12 alleles are risk factors for HCC in the whole populations, especially in the Asians; DRB1*15 allele is only associated with an increased risk of HCC in Asians (under the fixed effect model). However, it is necessary to conduct large trials using standardized unbiased methods, homogeneous HCC patients and well matched controls, with the assessors blinded to the data. Moreover, gene-gene and gene-environment interactions should also be considered in the analysis. Such studies taking these factors into account may eventually lead to a better, more comprehensive understanding of the association between HLA-DRB1 polymorphism and HCC.HLA: human leukocyte antigens; HCC: hepatocellular carcinoma; HBV: hepatitis B virus; OR: odds ratio; CI: confidence interval.The authors declare that they have no competing interests.ZHL and YNX carried out the design of this meta-analysis, conducted a searching, extracted data, analyzed the data and drafted the manuscript. SYX participated in study design and the critical revision of the manuscript. QJD, QW, SHZ, XJJ and YS participated in the critical revision of 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/1471-230X/10/145/prepubPRISMA 2009 Checklist. We conformed to PRISMA statement standards and provided the detailed information.Click here for file


MicroRNAs (miRNAs) participate in various cellular processes such as cell growth, differentiation, cell death and play an important role in a variety of diseases, especially in cancer. Recently, a number of studies have investigated the association between single nucleotide polymorphisms (SNPs) on the hsa-miR-149 rs2292832 and susceptibility to cancer; however, the results remain inconclusive.We carried out a meta-analysis of 12 studies including 5937 cases and 6081 controls from PubMed to assess the association between the hsa-miR-149 rs2292832 and cancer risk by pooled odds ratios (ORs) and 95% confidence intervals (CIs). However, our results showed that genotype distribution of the hsa-miR-149 rs2292832 was not associated with cancer risk in all genetic models. Subgroup analysis by cancer type, ethnicity or study design showed no significant association either.Results of this meta-analysis suggest that the hsa-miR-149 rs2292832 polymorphism is not associated with cancer risk in spite of the potentially protective role of C allele in hepatocellular carcinoma and male gastric cancer. MicroRNAs (miRNAs) are an abundant class of small non-coding RNAs that negatively regulate gene expression by base pairing with the 3’-untranslated region of target mRNAs, resulting in either mRNA cleavage or translational repression ,2. Many Single nucleotide polymorphisms (SNPs) located at miRNA genes may be associated with changes in miRNA processing, thus leading to functional changes of miRNA by influencing interaction between miRNAs and their target mRNAs ,10. ManyRecently, many studies have investigated the association between the hsa-miR-149 rs2292832 polymorphism and cancer risk. But the results were not conclusive and consistent. Considering the limits of the single study, we performed this meta-analysis of 12 published studies to derive a more powerful estimation of the association between the hsa-miR-149 rs2292832 polymorphism and cancer risk.Medical subheading (Mesh) terms: ‘microRNA’, ‘cancer’ and ‘polymorphism’ were used to search on PubMed for eligible studies . The references of articles and reviews were also examined to explore potentially additional studies. Studies were eligible if they met the following criteria: (a) case-control studies; (b) investigating the association between the hsa-miR-149 rs2292832 polymorphism and cancer risk; (c) detailed genotype data for estimating of odds ratio (OR) and 95% confidence interval (CI); (d) full text articles in English. If multiple studies had overlapping or duplicate data, only those with complete data or recent studies were included.Data were evaluated and extracted from the eligible studies by two investigators (Xu and Zhou) independently. The following items from each study were recorded: first author’s name, year of publication, country or area of origin, ethnicity, cancer type, source of controls, genotyping method, total number of cases and controls, genotype distributions of cases and controls, and Hardy-Winberg equilibrium (HWE), respectively. If discrepancies existed between two investigators, another investigator (Qiu) was invited to discuss and check the data until a consensus was reached.CT/CC), dominant model (TT/CT vs. CC), homozygote comparison (TT vs. CC), and heterozygote comparison (CT vs. CC), respectively. The statistic significance of pooled ORs was determined by Z-test and a p<0.05 was considered as statistically significant. A chi-square based Q-test was used to check the heterogeneity among the studies. A p<0.10 for Q-test suggested significant heterogeneity among studies, and the random-effects model (DerSimonian-Laird method) was conducted to calculate the pooled ORs [HWE was evaluated for controls in each study by the chi-square test and a p<0.05 was considered as departure from HWE. Odds ratios (ORs) with 95% confidence intervals (CIs) were used to assess the strength of association between the hsa-miR-149 rs2292832 polymorphism and cancer risk. Pooled ORs were performed for allelic comparison (T vs. C), recessive model  . Thus, a trim and fill method was used and pooled ORs were recalculated with hypothetically non-published to evaluate the asymmetry in the funnel plot , indicating the stability of the results.We used Begg’s funnel plot and the Egger’s linear regression test to assess publication bias. Publication bias was detected Click here for additional data file.Table S1PRISMA checklist.(DOC)Click here for additional data file.


Gliomas account for almost 80% of primary malignant brain tumors. Epidermal growth factor (EGF) is an interesting research candidate in which to look for genetic polymorphisms because of its role in mitogenesis and proliferation. Extensive studies have found that a single nucleotide polymorphism (SNP) +61G/A (rs4444903) in the EGF gene is associated with the susceptibility of glioma, however, the results have been controversial. Furthermore, the association between EGF +61G/A polymorphism with the development and grade progress of glioma has not been established.We examined the association of EGF +61G/A polymorphism and glioma by performing a meta-analysis. Nine studies testing the associations between EGF +61G/A polymorphism and risk of glioma with 1758 cases and 2823 controls were retrieved. Odds ratios (ORs) and 95% confidence intervals (CIs) were used to assess the strength of the association. The pooled ORs were performed for the allele model, codominant model, dominant model, and recessive model, respectively.Overall, this meta-analysis showed significant associations between the EGF +61G/A polymorphism and glioma susceptibility in all four genetic models. However, in the stratified analysis by the grade of glioma, we only found this association existed in patients with Grade IV glioblastoma, but not in patients with Grade I-III glioma. We further compared EGF +61G/A polymorphism in patients with glioblastoma and Grade I-III glioma accordingly, the stronger association between the EGF +61G/A polymorphism and the malignancy of glioma was found.The results of this meta-analysis suggested that the EGF +61G/A polymorphism is associated with both the susceptibility of glioma and the malignance of glioma. Gliomas are the most common primary tumors in the central nervous system The epidermal growth factor (EGF) gene, which is a member of the EGF superfamily, has been demonstrated to activate cell proliferation and stimulate mitogenesis in epidermal tissues To date, several studies have found the association between the EGF +61G/A polymorphism with the susceptibility of glioma, however, these studies showed inconsistent results The procedures of this meta-analysis adhered to the PRISMA Statement guidelines We did a literature search in online database such as PubMed, Embase and the Chinese knowledge library using the search terms “epidermal growth factor”, “polymorphism”, and “glioma”, along with additional terms such as “EGF, SNP, Glioblastoma, astrocytoma, oligodendroglioma”, and all possible combinations . The search was limited to English-language articles. In order to search more deeply, we reviewed the references of the selected articles to retrieve data that we could have ignored in the initial search. Of the studies with overlapping data published by the same researchers, we selected the most recent ones with the largest number of subjects.The studies for the meta-analysis were selected when they satisfied the following criteria: studies about EGF +61G/A polymorphism and glioma risk, case-control studies in humans, studies containing useful genotype frequencies or odds ratio (OR) data, studies containing classification of glioma grade. Major exclusion criteria were (a) no control population, (b) no available genotype frequency, (c) no explicit glioma grade classification, and (d) duplication of the previous publication.Information was extracted from all eligible articles independently by two of the authors following the criteria listed above (Xin Chen and Guang Yang). Disagreement was discussed and resolved between the two authors.For each study, the following data were considered: the first author's last name, year of publication, country of origin, ethnicity and numbers of genotyped cases and controls. We did not define any minimum number of patients to include a study in our meta-analysis. In the event that a study presented subpopulations, these were taken to be different studies.The estimate of association between the EGF +61G/A polymorphism and glioma risk was assessed using the fixed-effect method The analysis of heterogeneity between studies was performed by the Q statistic, with p-values<0.05 indicating significant heterogeneity Among the 9 publications, there were five studies of Caucasian descendants The evaluation of association between EGF +61G/A polymorphism and glioma risk are presented in At the same time, on the basis of glioma grade subgroup, we further compared the EGF +61G/A polymorphism in patients with glioblastoma and Grade I-III glioma, and found the OR value was greater than 1 , in the sense that by contrast to Grade I-III glioma, glioblastoma has the stronger associations with EGF +61G/A polymorphism .To assess the influence of each individual study on the pooled ORs, we performed a sensitivity analysis by omitting individual study. The analysis results suggested that no individual study significantly affected the pooled ORs under the allele model (a versus A) of EGF +61G/A polymorphism , which iBegg's funnel plot and Egger's test were performed to assess the publication bias of included literatures. The shapes of the funnel plots in all genetic models of EGF +61G/A polymorphism seemed symmetrical , suggestCell growth dysregulation is one important characteristic of tumors. EGF/EGFR signaling dysregulation often occurs in gliomas and plays a key role in gliomagenesis Many retrospective studies have shown that polymorphism in EGF or EGFR genes strongly correlates with susceptibility to glioma. Some studies have reported the association between polymorphism in EGF +61G/A and glioma susceptibility From the above, we concluded that EGF +61G/A polymorphism is associated with the susceptibility to glioma. As the literature reported, EGF +61G/A polymorphism is also probably associated with the survival time of glioma patients Our approach allowed us to look for potential glioma grade differences in the association between EGF +61G/A polymorphism and glioma risk. Analysis of glioma grade subgroups showed a strong association between subjects carrying at least one “a” allele with the risk of glioblastoma (Grade IV), but this association was not found in Grade I-III glioma. We further compared the EGF +61G/A polymorphism in patients with glioblastoma and Grade I-III glioma, and found that glioblastoma has a stronger association with EGF +61G/A polymorphism compared with Grade I-III glioma, this means there is a significant association between the EGF +61G/A polymorphism and the malignancy of glioma.We think this result will strengthen the previous knowledge about EGF +61G/A polymorphism and glioma risk. This study has an important implication for glioma progress prediction that genotyping for the EGF +61G/A variant might help in identifying patients at high risk of glioblastoma. However, it needs to be emphasized that our meta-analysis involved only one study with caucasian Grade I-III glioma population, so this association should be re-evaluated in studies with larger sample sizes. Therefore, the results in the present study should be interpreted with caution.There are some limitations that should be considered to interpret this result. Confounding factors may affect the meta-analysis. Firstly, detailed information such as age and gender in cases and controls of different genotypes in some studies were not available, which limited further analysis. Secondly, the numbers of published studies were not enough for a comprehensive analysis, particularly for different grades of glioma, we could not obtain enough statistical power to investigate the real association. Because of the limited analysing data, we can only conclude that EGF +61G/A polymorphism is related to high grade gliomas. In current results, the correlativity between EGF +61G/A polymorphism and low grade glioma is not clear. Until now, little is known about its function in primary and secondary glioblastomas. As Kunihiko Watanabe reported that immunoreactivity for the EGFR prevailed in primary glioblastomas but was rare in secondary glioblastomas Despite of these limitations, our meta-analysis still has clear advantages over previous studies. To the best of our knowledge, this is the first meta-analysis of the association between EGF +61G/A gene polymorphisms and the susceptibility to glioma subject to different grades. In addition, data were abstracted from different studies, which significantly increased statistic power of this meta-analysis. Meanwhile, studies included in this meta-analysis were case–control studies and contained available genotype frequency, which met our selection criterions.In summary, this meta-analysis suggested that the EGF +61G/A polymorphism is associated with glioma development risk. Additionally, we found that this association was more prominent in glioblastoma than in Grade I-III glioma, showing that the EGF +61G/A polymorphism is not only associated with the susceptibility to glioma, but also with the malignance of glioma. Future well designed larger studies are warranted to validate this association in different grade populations of glioma.Figure S1Forest plot of glioma risk for EGF +61G/A polymorphism (aa/Aa versus AA). The squares and horizontal lines correspond to the study-specific OR and 95% CI. The area of the squares reflects the study-specific weight (inverse of the variance). The diamond represents the pooled OR and 95% CI.(TIF)Click here for additional data file.Figure S2Forest plot of glioma risk for EGF +61G/A polymorphism (aa/Aa versus AA) by glioma grade. The dots and horizontal lines correspond to the study-specific OR and 95% CI. The diamond represents the pooled OR and 95% CI.(TIF)Click here for additional data file.Figure S3Forest plot of EGF +61G/A polymorphism associated with glioblastoma and Grade I-III glioma risk (aa/Aa versus AA). The squares and horizontal lines correspond to the study-specific OR and 95% CI. The area of the squares reflects the study-specific weight (inverse of the variance). The diamond represents the pooled OR and 95% CI.(TIF)Click here for additional data file.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


Glutathione S-transferases M1 (GSTM1) is an important phase II metabolizing enzyme. The null genotype of GSTM1 causes total loss of GSTM1 enzyme activity and numerous studies have investigated the association between GSTM1 null genotype and gastric cancer risk.This meta-analysis was designed to investigate the relationship between GSTM1 null genotype and susceptibility to gastric cancer and assess the influence of Helicobacter pylori infection, smoking, Lauren’s classification, and other factors. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to estimate the association strength.heterogeneity<0.001). Sub-group analysis suggested that the significant association was only observed in Asians , but not in Caucasians. The increased risk was found among H. pylori positive population , while no association was found among H. pylori negative population . For smoking status, the GSTM1 null genotype increased risk of gastric cancer in both ever-smokers and non-smokers. Source of control, sample size, location of tumor and Lauren’s classification did not modify the association.A total of 46 eligible studies were indentified and analyzed in this meta-analysis, including 8138 cases of gastric cancer and 13867 controls. Pooled results showed that the GSTM1 null genotype was associated with a significantly increased risk of gastric cancer (OR=1.217, 95% CI: 1.113-1.331, PIn this meta-analysis based on 46 epidemiological studies, we show that the GSTM1 null genotype is associated with an increased risk of gastric cancer among Asians but not among Caucasians. H. pylori infection but not smoking status could modify the association. It has been well demonstrated that gastric cancer is the second leading cause of cancer related death and the fourth most common malignancy, which accounts for 9.7% of total cancer deaths worldwide . As a maHuman glutathione S-transferases (GSTs) are phase II metabolizing enzymes that are critical for protection from cancer by detoxifying numerous potentially cytotoxic or genotoxic compounds . AccordiA lot of epidemiological studies have investigated the association of GSTM1 depletion with risk of gastric cancer and several meta-analyses have been performed to clarify this issue ,8,9. TheThus, we conducted an updated meta-analysis to comprehensively assess the relationship between GSTM1 deletion polymorphism and risk of gastric cancer and evaluate the influence of confounding factorsThis study was carried out and reported in agreement with the PRISMA guidelines for systematic reviews and meta-analyses  (supplemStudies were selected according to the following inclusion criteria: (1) case-control studies; (2) investigating the association between GSTM1 deletion polymorphism and gastric cancer risk; (3) with available genotype distribution data to calculate combined odds ratios (ORs) and 95% confidence intervals (95% CIs). Studies without detail genotype distribution data were excluded. Titles and abstracts of searching records were primarily screened and full text papers were further retrieved to confirm eligibility. Two reviewers (ZY and DX) extracted eligible studies independently according to the inclusion criteria. Disagreement between two reviewers was discussed with another reviewer (SGQ) till consensus was achieved.Data of eligible studies was extracted by two reviewers (ZY and DX) independently with a pre-designed data-collection form. The following data was collected: name of first author, year of publication, country where the study was conducted, ethnicity, source of control, number of cases and controls, genotype frequency in cases and controls, H. pylori infection (positive and negative), smoking status (ever-smoker and non-smoker), tumor location (cardia and non-cardia), and Lauren’s classification . Different ethnicity descents were categorized as Asian, Caucasian, and Latin American. For H. pylori infection and smoking status, we collected data according the original definition of eligible studies and no modification or adjustment was performed. Eligible studies were defined as hospital-based (HB) and population-based (PB) according to the control source. Two reviewers reached consensus on each item.The association strength between GSTM1 null genotype and gastric cancer risk was measured by OR with 95% CI. The estimated ORs were achieved by pooling genotype distribution data from each eligible study. A 95% CI was used for statistical significance test and a 95% CI without 1 for OR indicating a significant increased or reduced cancer risk. The pooled ORs were calculated for the comparison of null genotype versus present genotype. Subgroup analyses were also conducted to explore the effects of confounding factors: ethnicities, sources of control, sample size, H. pylori infection, smoking status, location, and Lauren’s classification. Chi-square based Q test was used to check the statistical heterogeneity between studies, and the heterogeneity was considered significant when p<0.10. Given t0.05) and p=0.01 (Nfs0.01) was also calculated[Publication bias was detected with Begg’s funnel plot and the Egger’ linear regression test, and a p < 0.05 was considered significant. To testalculated.All statThe detailed process of study selection was shown in heterogeneity<0.001; 0.05=1299.9, Nfs0.01=602.8), which suggested that the publication bias was feeble and our result is solid.All meta-analysis results were shown in heterogeneity=0.002), while no significant association was found in Caucasians or Latin Americans , while no significant association was found in the HP negative sub-group .HP infection is a well known risk factor of gastric cancer and 4 studies provided data about HP infection status and GSTM1 genotype distribution. As shown in Smoking is a risk factor of various kinds of cancer, including gastric cancer, and GST family is also involved in the metabolism of various carcinogens in cigarette smoke. As shown in We also performed stratified analyses according to location of tumor (cardia and non-cardia) and Lauren’s classification . The number of studies available for the sub-group analysis of location was quite small (3 studies), and no significant association of GSTM1 null genotype with gastric cancer risk was observed in neither sub-group . As for Genetic polymorphisms are natural DNA sequence variations and the expected frequency is about 1% among healthy population . FunctioSince the first study in 1991 by Strange and colleagues  which reIn this study, we identified 46 eligible studies, including 8138 gastric cancer cases and 13867 controls, which could provide sufficient statistic power. By pooling all available data, we found the null genotype was associated with a statistically elevated risk of gastric cancer , which was consistent with previous meta-analyses . By straHelicobacter pylori, the group I carcinogen classified by World Health Organization, is one of the most important risk factors for gastric cancer. By perf0.05=1299.9, Nfs0.01=602.8) to provide credence to our results. Secondly, heterogeneity was significant in this study. To achieve a precise and conservative estimation, we used random-effects model to pool eligible studies and meta-regression found that ethnicities, source of control, and sample size were the source of heterogeneity. Thirdly, in the sub-group analysis of location, number of studies was relatively small and the results should be interpreted with caution.Compared with previous meta-analysis, we included more studies and performed sub-group analyses to assess the influence of ethnicities, source of controls, sample size, H. pylori infection, smoking, tumor location, and Lauren’s classification. Notably, we searched databases of PubMed and EMBASE but not China National Knowledge Infrastructure (CNKI), because CNKI Chinese-language database, which is usually not accessible for non-Chinese researchers. However, limitations of this meta-analysis should be highlighted. First, Egger’s test and Begg’s test suggested the evidence of publication bias. We calculated the fail-safe number and the number was large enough Click here for additional data file.Figure S1Sub-group analysis of ethnicities.(TIF)Click here for additional data file.


PRNP). Breeding programs have been implemented to increase scrapie resistance in sheep populations; though desirable, a similar approach has not yet been applied in goats. European studies have now suggested that several polymorphisms can modulate scrapie susceptibility in goats: in particular, PRNP variant K222 has been associated with resistance in case-control studies in Italy, France and Greece. In this study we investigated the resistance conferred by this variant using a natural Italian goat scrapie isolate to intracerebrally challenge five goats carrying genotype Q/Q 222 (wild type) and five goats carrying genotype Q/K 222. By the end of the study, all five Q/Q 222 goats had died of scrapie after a mean incubation period of 19 months; one of the five Q/K 222 goats died after 24 months, while the other four were alive and apparently healthy up to the end of the study at 4.5 years post-challenge. All five of these animals were found to be scrapie negative. Statistical analysis showed that the probability of survival of the Q/K 222 goats versus the Q/Q 222 goats was significantly higher (p = 0.002). Our study shows that PRNP gene mutation K222 is strongly associated with resistance to classical scrapie also in experimental conditions, making it a potentially positive target for selection in the frame of breeding programs for resistance to classical scrapie in goats.Susceptibility of sheep to scrapie, a transmissible spongiform encephalopathy of small ruminants, is strongly influenced by polymorphisms of the prion protein gene ( Sc) of a host-encoded cellular prion protein (PrPC) ) di-, mono- and un-glycosylated band intensity was 48.16 ± 2: 31.86 ± 2: 19.98 ± 1.9, respectively, for the inoculated goats and 49.27 ± 1.5: 32.19 ± 1.1: 18.54 ± 2, respectively, for the inoculum sample. Quantification showed that the PrPSc signal intensity in the positive muscular tissues was generally similar to that obtained when five milligrams of the corresponding positive brainstem homogenate were diluted in five grams of a negative muscle homogenate. From this finding, we estimated that the PrPSc levels in the three examined cases were lower than those found in the corresponding brainstems by a factor of approximately 1 × 10-3. The signal intensities of the positive omasum, abomasum and kidney were usually lower than those found in the corresponding brainstems by a factor of 1 × 10-3 - 1 × 10-4.All five Q/Q 222 animals tested positive for the CNS and the large part of the lymph reticular system analysed both by Western blot and immunohistochemistry. Moreover, the peripheral nerves and numerous extraneural organs tested scrapie positive by Western blot Table . MoleculThe Q/K 222 goats were completely negative by immunoblot analysis of all extraneural tissues and organs.PRNP gene mutation K222 is strongly associated with resistance to classical scrapie in experimentally challenged goats, confirming the results previously obtained in field studies [Our study shows that  studies ,21. The  studies . Similar studies  reported studies . HoweverPRNP gene, but it was the only one out of its group that was homozygote for serine at codon 240. While this prolonged i.p. in one animal does not allow us to say whether it was significant or just happened by chance, we can speculate that other unknown genetic factors may have come into play or that indeed serine at codon 240 could have had an effect. Field studies have produced controversial results about codon 240, indicating a positive association with scrapie infection of P240 [The results of the challenge of the Q/Q 222 goats yielded further interesting information on goat scrapie genetics. The two goats carrying the octarepeat deletion died of scrapie, with an i.p. not unlike that of the other goats in the same group, showing that in our sample this mutation did not confer any resistance, on the contrary to that which was hypothesized by Goldmann , perhaps of P240  or a pos of P240 ,16,17. I of P240  reported of P240  found thThe clinical signs seen in three of the five scrapie positive animals were similar to those previously described in both experimental and natural scrapie, but none of the animals in the present study showed aggressiveness, hypersensitivity to external stimuli or hyperexcitability. As already known in goats, unlike that observed in sheep with scrapie, ,33-35, sSc distribution in the extraneural tissues; nonetheless, it does give information on the centrifugal diffusion of PrPSc, which may prove useful for risk assessment of human exposure given by peripheral districts. Our results are in line with previous observations in goats and sheep with natural scrapie or after oral challenge. It appears that PrPSc had a widespread distribution, similar to that observed in sheep, involving the lymph reticular system, as already known in goats [Sc deposition in sheep [Since the route of inoculation differed from that of the natural disease, no conclusions on the pathogenesis of scrapie in goats can be drawn from the pattern of PrPin goats , and othin sheep ,38,28. Sin sheep , scrapiein sheep ,41.Sc amount, compared to that in the brainstem, was at least 1000-fold lower in the muscles and 1000 to 10 000-fold lower in the omasum, abomasum and kidney, suggesting a low risk for human transmission by these tissues.The finding of positive signals in the skeletal muscles in three of the five scrapie-positive goats is consistent with reports of muscle tissues that tested positive in a scrapie i.c. challenged goat  and oralPRNP gene can be a genetic target to select for in the frame of breeding programs for the control and eradication of classical scrapie in goats. This result is certainly mainly valid for Italy, given that an Italian scrapie isolate was used in the experiment. The situation in Italy could be particularly favorable because there is good evidence for a low variability of circulating scrapie strains [After identifying a genetic variant associated with TSE resistance, the main question is the validity of the association in relation to different classical scrapie strains, atypical scrapie and BSE. Our study confirms that variant K222 of the goat  strains ,45. The  strains . On the  strains . No data strains . Similar strains .One limitation of the present study was the lack of homozygous K/K222 animals. The role of homozygotes needs to be investigated to verify whether there could be a situation like that in sheep, where ARR/ARR animals are highly resistant, while some cases of disease are occasionally reported in ARR/XXX . BarilleAnother open question in the present study is whether negative K222 goats can be healthy carriers of infection. To answer this, the negative tissues of the challenged Q/K 222 goats will be tested by bioassay for the presence of infectivity.Other experimental challenge studies on the K222 variant are underway in Europe . If theyThe authors declare that they have no competing interests.Sc quantification, and contributed to the manuscript; ADA carried out the neurological clinical evaluations, interpreted the clinics, and contributed to the manuscript; SP assisted in observation of the animals, sample collection, and interpreting the results of genetics; SC carried out the genetic analysis to find the animals and kept the database and records; CM performed the statistical analysis and contributed to the manuscript; CP carried out the immunohistochemistry; BI assisted in sample collection and interpreting the immunohistochemistry results; MM interpreted the immunoblot results; LDA carried out the immunoblot analysis; FZ assisted in observation of the animals; CC prepared the inoculum and assisted in sample collection; NM managed daily observation of the animals and assisted in sample collection; CC interpreted the immunohistochemistry results; MC contributed to the manuscript; GL inoculated the animals and was responsible for their care. All the authors read and approved the final manuscript.PLA conceived and led the project, interpreted the genetics and drafted the manuscript; FM prepared the inoculum, assisted in sample collection, performed PrP