Calcium nephrolithiasis may be considered as a complex disease having multiple pathogenetic mechanisms and characterized by various clinical manifestations. Both genetic and environmental factors may increase susceptibility to calcium stones; therefore, it is crucial to characterize the patient phenotype to distinguish homogeneous groups of stone formers. Family and twin studies have shown that the stone transmission pattern is not mendelian, but complex and polygenic. In these studies, heritability of calcium stones was calculated around 50%Calcium-sensing receptor (CaSR) is mostly expressed in the parathyroid glands and in renal tubules. It regulates the PTH secretion according to the serum calcium concentration. In the kidney, it modulates electrolyte and water excretion regulating the function of different tubular segments. In particular, CaSR reduces passive and active calcium reabsorption in distal tubules, increases phosphate reabsorption in proximal tubules and stimulates proton and water excretion in collecting ducts. Therefore, it is a candidate gene for calcium nephrolithiasis.In a case-control study we found an association between the normocitraturic stone formers and two SNPs of CaSR, located near the promoters region (rs7652589 and rs1501899). This result was replicated in patients with primary hyperparathyroidism, comparing patients with or without kidney stones. Bioinformatic analysis suggested that the minor alleles at these polymorphisms were able to modify the binding sites of specific transcription factors and, consequently, CaSR expression.Our studies suggest that CaSR is one of the candidate genes explaining individual predisposition to calcium nephrolithiasis. Stone formation may be favored by an altered CaSR expression in kidney medulla involving the normal balance among calcium, phosphate, protons and water excretion. According to modern criteria, calcium nephrolithiasis can be considered as a multifactorial disorder, including different groups of patients, homogeneous for their final phenotype but heterogeneous for their intermediate phenotypes and pathogenetic mechanisms. Therefore, it appears as a complex disease that may develop changeable clinical manifestations in urine composition, bone involvement, stone composition, age at onset and other intermediate phenotypes ,2. Both Studies on families and twins showed the genetic determination of nephrolithiasis. Stone formers were more common among the first-degree relatives of stone-forming patients than healthy individuals. Concordance for stones was greater among monozygous than dizygous twins. In both studies, hereditary of calcium stones, defined as the proportion of the phenotipic variance depending on genes, was approximately 50% -6.Family analysis also showed that the transmission pattern of calcium stones is not mendelian, but complex and polygenic, although 3 decades ago the first studies proposed an autosomal dominant hereditary transmission . The patIn our preliminary clinical approach to the disease, we evaluated the association of calcium nephrolithiasis with the polymorphisms of 10 candidate genes, selected on the basis of the knowledge of the possible role of their products in stone formation.We selected genes of two calcium pumps of the plasma membrane (ATP2B2 and ATPB1), the anion channel (SLC4A1), the calcium-sensing receptor (CaSR), a phosphate carrier (SLC34A1), the vitamin D receptor (VDR), two dicarboxylic acid carriers (SLC13A1 and SLC13A2), the lysosomial H-pump (ATP6N1B), the calcium channel of the distal tubule (TRPV5). We chose 2-6 biallelic polymorphisms on each gene (Unpublished data). We only found an association between calcium stones and rs1501899 (G > A), a SNP located on the first intron of the CaSR gene. The minor allele A at rs1501899 was more frequent in stone formers than healthy subjects . It. It25]. CaSR is expressed on the basolateral membrane of the thick ascending limb of Henle loop . In thisq proteins [In the distal convoluted tubule, CaSR is located on the basolateral membrane of tubular cells and in cultured dog cells from this tubular segment CaSR was found to reduce the active calcium reabsorption by interfering with the calcium pump function through a phospholipase C dependent mechanism. The signaling pathway for this effect requires activation of Gproteins .In the collecting duct, CaSR is expressed on the apical membrane of the principal and intercalated cells ,31. In cThe analysis of the described CaSR functions in the kidney suggests that in the ascending limb and distal convolute tubule CaSR is sensitive to serum calcium because of its location on basolateral membrane of tubular cells. Here, the CaSR modulates calcium reabsorption according to the serum levels of calcium so that its increase may be compensated by the CaSR-mediated inactivation of both passive and active distal calcium reabsorption. Nevertheless, a high calcium excretion is potentially dangerous for the kidney as it increases the probability of calcium-phosphate precipitation inside the kidney. This probability especially increases in the papilla, where the fluid concentration is unusually high. To prevent possible dangerous consequences caused by high calcium excretion, CaSR might induce the modifications counterbalancing the tubular function. In proximal tubules, the antiphosphaturic effect reduces the phosphate load to the distal tubular segments, where CaSR decreases calcium reabsorption. In the collecting duct, CaSR promotes urinary salt dilution and acidification, both necessary to favor the solubility of calcium-phosphate salts . These eThe relevance of CaSR in calcium homeostasis and our preliminary findings about stone former genotyping led us to study more in depth the possible role of CaSR gene polymorphisms in calcium kidney stones formation. For this purpose, we selected stone forming patients presenting a normal excretion of citrate, because the pathogenetic effect leading to calcium-phosphate precipitation may be predominant compared to the function performed by the precipitation inhibitors, among which urine citrate is one. On the contrary, urine calcium and phosphate promote salt precipitation and CaSR, influencing their tubular handling, may affect urine saturation for their salts .CaSR gene at a distance of 5-10 kb [Our study included 463 patients affected by both calcium phosphate and oxalate stone disease of which 312 were normocitraturic and 151 hypocitraturic, and 213 healthy subjects. All of them were genotyped for 21 biallelic polymorphisms mapping the whole  5-10 kb . The fre 5-10 kb .In order to confirm this study, we replicated it on a second sample of patients presenting primary hyperparathyroidism (PHPT). These patients frequently suffer from stone disease, as they are hypercalciuric  and normocitraturic. This study compared 157 PHPT patients forming kidney stones and 175 PHPT patients not forming stones. Also in this population polymorphisms rs7652589 and rs1501899 were associated with the presence of stones .Polymorphisms rs7652589 and rs1501899 are respectively at 11000 kb before P1 and at 4600 kb after P2 figure . This loFurther genotype analysis in normocitraturic patients showed that calcium stones were associated with polymorphisms located within P1, whereas P2 polymorphisms were not associated . These fThe hypothesis we are proposing is based on association findings and has to be supported by functional data that, confirming it, could clarify the influence of the polymorphisms on the gain or loss of the CaSR expression.Some data indicate that minor alleles at the associated polymorphism lead to a loss of CaSR expression in the kidney. Supporting this finding, serum levels of PTH were lower in normocitraturic stone formers than in homozygous patients for the variant allele at rs7652589 and rs1501899 . These fThree missense polymorphisms of the CaSR gene have a significant frequency in general population. They are located on exon 7 and are A986S , R990G  and Q1011E . Few studies considering CaSR as a candidate-gene tested the frequency of their alleles in nephrolithiasis. A study was performed in 223 Canadian idiopatic calcium stone formers genotyped for A986S and R990G. It observed significant Hardy-Weinberg disequilibrium at the R990G locus in cases, but not in controls, and attributed this disequilibrium to the association of the polymorphism with stones. Carriers of the 990G allele had an eightfold increase of stone risk . AnotherCaSR gene polymorphisms may be involved in calcium nephrolithiasis, but propose an intriguing scenario in which stone formation appears to be favored by an activating polymorphism (R990G) and by polymorphisms decreasing expression of the CaSR gene. Despite their apparently opposite effect, both polymorphisms could predispose to calcium nephrolithiasis. Today, we have no sufficient knowledge to explain this puzzle, that, to be resolved, needs experiments evaluating the CaSR activity in kidney tubules and the functional effect of the variant alleles at these polymorphisms. An initial contribution was given by a study in knockout mice for the calcium channel TRPV5 [These findings confirm that el TRPV5 . TRPV5-/el TRPV5 .Our studies suggest that CaSR is a candidate gene to explain individual predisposition to calcium kidney stones . Stone fMore extensively, our findings indicate that CaSR could prevent deposition of calcium-phosphate salts in tissues. Immunohistochemistry studies and RT-PCR showed that CaSR expression is decreased in calcified arteries from both uremic and non-uremic patients ,49. StudThese data suggest a physiopathological role of CaSR in protecting human tissues against the risk of calcification, a risk induced by the condition of supersaturation in body fluids. The functional importance of CaSR is highlighted by the fact that a CaSR precursor is detectable in unicellular organisms where it senses environmental salinity. The evolution of this molecule provided human beings with a cellular calcium sensor used by kidney and parathyroid cells to keep serum calcium within the normal range. CaSR anticalcification function becomes more complex in human kidney where it contributes to the paracrine regulation of ions and water excretion, thus protecting against calcium-phosphate precipitation. Its anticalcification activity has been hypothesized also in arteries but its mechanism has still to be ascertained and understood.The authors declare that they have no competing interests.All authors have read and approved the final manuscript.GV, AT and LS drafted the manuscript. LS and AT designed the experiment, drafted the experiments and troubleshooting. AT, ED and AA were responsible for the laboratory assay and performing the experiments. GV, MC, FR, DC and TA participated in clinical study.GV MD, nephrologist, Principal Investigator of Clinical StaffAT PhD in Molecular MedicineFR MD, Fellow of NephrologyTA MD, nephrologistMC MD PhD, nephrologistAA biologist, PhD student in Molecular MedicineED PhD, NutritionistDC MD, nephrologistLS PhD, Principal Investigator of Experimental Staff


Late-stage age-related macular degeneration (AMD) is a common sight-threatening disease of the central retina affecting approximately 1 in 30 Caucasians. Besides age and smoking, genetic variants from several gene loci have reproducibly been associated with this condition and likely explain a large proportion of disease. Here, we developed a genetic risk score (GRS) for AMD based on 13 risk variants from eight gene loci. The model exhibited good discriminative accuracy, area-under-curve (AUC) of the receiver-operating characteristic of 0.820, which was confirmed in a cross-validation approach. Noteworthy, younger AMD patients aged below 75 had a significantly higher mean GRS  than patients aged 75 and above . Based on five equally sized GRS intervals, we present a risk classification with a relative AMD risk of 64.0 (95% CI: 14.11–1131.96) for individuals in the highest category  compared to subjects with the most common genetic background . The highest GRS category identifies AMD patients with a sensitivity of 7.9% and a specificity of 99.9% when compared to the four lower categories. Modeling a general population around 85 years of age, 87.4% of individuals in the highest GRS category would be expected to develop AMD by that age. In contrast, only 2.2% of individuals in the two lowest GRS categories which represent almost 50% of the general population are expected to manifest AMD. Our findings underscore the large proportion of AMD cases explained by genetics particularly for younger AMD patients. The five-category risk classification could be useful for therapeutic stratification or for diagnostic testing purposes once preventive treatment is available. Age-related macular degeneration (AMD) is a common degenerative disease of the central retina and a leading cause of severe vision impairment in Western societies CFH) on 1q32 and age-related maculopathy susceptibility 2 (ARMS2)/HtrA serine peptidase 1 (HTRA1) on 10q26 CFH locus indicate at least three independent risk variants ARMS2/HTRA1 region, a single risk haplotype was found to fully explain the observed association AMD is a complex disease influenced by genetic and environmental factors with estimates of heritability varying from 45% to 71% C2)/complement factor B (CFB) C3) CFI) LIPC) on chromosome 15q22 CETP) and the lipoprotein lipase precursor (LPL) genes APOE) gene SYN3) and approximately 100 kb upstream of the tissue inhibitor of metalloproteinases-3 gene (TIMP3) A crucial role of the complement system in AMD pathogenesis was further supported by subsequent candidate gene studies. These studies identified risk-associated variants in or near three additional complement genes including the complement component 2 . There were three further SNPs with reportedly established association, which we did not select for the model: (i) at the CFH locus, an association of four variants with AMD is known ; however, rs1410996 is present on two distinct haplotypes, each of which is tagged by rs800292 (correlation r2 = 0.473 to rs1410996 2 = 0.283 to rs1410996 2 = 0.008 2 = 1), rs10490924 was reported to fully capture the disease risk at this locus Eight loci  with 13 SNPs and established association with AMD were included into our genetic risk score modeling  in 986 cases and 796 controls from the Lower Frankonian AMD case-control study Table 1..For each SNP, association with AMD was computed using a logistic regression model, unadjusted for age or gender Table 2.Based on the data from the 13 SNPs, we fit a multiple logistic regression model Figure 1Computing the area-under-the-curve (AUC) of the receiver-operating characteristic for the 13-SNP GRS, we observed good ability to correctly classify those with and without the disease  random sub-sampling with 2/3rd of the data for model building and 1/3rd for testing. The cross validated AUC of 0.813 (95% CI: 0.813–0.814) is close to the one described in our initial study (AUC = 0.820).2 = 0.247)  and equal model fit (R= 0.247) Table 3.The distribution of GRS for cases and controls as observed in our study is given in We further investigated differences of the GRS between age-groups (below or older than 75 years), men and women, or types of AMD  using a significance level of 0.05/3 to account for the three subgroup tests performed.−5), but there was no difference between the age-groups among controls (p = 0.18). The OR per GRS unit was 3.06 (95%CI: 2.64–3.59) for younger and 2.71 (95% CI: 2.44–3.05) for older individuals. We also found that the AUC restricted to the younger subjects (cases and controls) was higher (0.852) than when only older subjects (cases and controls) were included in the calculations (0.809).Significant differences in mean GRS were found between younger  and older  AMD cases (p = 8.7×10−5). It was also higher when compared to GA cases , although the latter was statistically not significant when applying a conservative Bonferroni-adjusted significance level of 0.05/3. The OR per GRS unit was also higher for mixed GA+NV cases  than for NV cases  or for GA cases . This effect appeared to be independent of age, since mean age in GA , NV  and mixed GA+NV  was similar. There was no significant difference in the GRS means between men and women neither among cases nor among controls.Cases with mixed GA+NV had a significantly higher mean GRS  compared to NV cases  when compared to NV or GA alone.To establish a classification scheme, we formed five equally sized intervals for the GRS spectrum –(−0.05), (−0.05)–1.70, 1.60–3.44, and >3.44; The relative risks were approximated as ORs for each GRS category using the middle category (no. 3) as reference Table 4.Due to the substantial differences found in mean GRS for younger compared to older cases (see above), we derived these ORs also separately by age-group. To avoid scarce data, risk group one and two as well as four and five were combined to a low and a high risk group, respectively Table 4.To reflect the anticipated situation in the general population and to compute the absolute risk of AMD per GRS group, we computed the fraction of late-stage AMD cases per GRS category by (i) utilizing the fraction of cases and controls as observed in each GRS category Table 4 Table S2).We again adopted the same cross-validation approach to compute absolute risks since the effect sizes of the variants in the GRS model, on which the absolute risk estimates are based, were estimated from our study data. This approach yielded overall similar estimates .). The slight discrepancy would be in-line with the fact that our controls were specifically selected to reveal no signs of early or late-stage AMD.As expected, the mean GRS was significantly higher in cases when compared to controls. Importantly, patients with late-stage AMD diagnosed at an earlier age had a significantly higher mean GRS than individuals that developed AMD later in life. This strongly suggests that genetic predisposition influences disease onset, which is also reflected in the higher relative AMD risk for younger subjects with an OR of 12.66 (95% CI: 6.76–25.65) when compared to older individuals with an OR of 5.18 (95% CI: 3.70–7.38). The mean genetic risk score in our control group was slightly lower but similar to the mean score in the HapMap sample . In a case-control setting, absolute risk or positive/negative predictive values cannot be derived without making assumptions on the overall AMD prevalence, which a prospective cohort study could estimate directly. Thus, the predictive ability of the risk score groups greatly depends on those assumptions. Second, it might be considered a limitation but also a strength that our study included exclusively late-stage AMD with NV or GA in one or both eyes as well as highly-matched controls with no signs of early or late-stage AMD in any eye. A strength as our data might exhibit less disease misclassification than other studies, but a limitation as the genetic relative risk could be overestimated if the genetic risk is larger for subjects with both eyes affected than for those with only one affected eye. Third, we had no independent and equally well characterized data set available to separate model building from testing although this is also the case for all other studies published on AMD risk score model building The highest genetic risk group of our proposed five-category classification scheme can effectively identify subjects at high risk for AMD. The specificity in this risk group was 99.9% (95% CI: 99.3%–100%). For example, our data and model suggest that 87.4% of subjects testing positive at some time in life for a high genetic risk are likely to develop AMD in their mid-eighties . Thus, this group of individuals could greatly profit from a sight-saving prevention or early intervention program while only 13% of  subjects would be alarmed and treated unnecessarily. However, still a large number of cases would be missed if this was established as a screening method (sensitivity 8.0% (95% CI: 6.5%–9.9%), i.e. 92% of all AMD cases would not be found in the highest risk group). Also individuals in the second highest risk group could possibly profit from early intervention, which would increase sensitivity to 47.6% and decrease specificity to 91.2%. However, this would only be acceptable, if the prevention/intervention is not harmful to the 59.9% of subjects treated and alarmed unnecessarily . These numbers are well in the range of established screening tests, e.g. for prostate cancer by prostate specific antigen (PSA) . Direct sequencing was performed with the Big Dye Terminator Cycle Sequencing Kit Version 1.1  according to the manufacturer's instructions. Reactions were analyzed with an ABI Prism Model 3130xl Sequencer (Applied Biosystems). TaqMan Pre-Designed SNP Genotyping Assays (Applied Biosystems) were performed according to the manufacturer's instructions. Additionally, some variants were genotyped by PCR followed by restriction enzyme digestion  and subsequent restriction fragment length analysis. The c.del443ins54 variant in the 3′-region of the ARMS2 locus was genotyped by a single PCR with oligonucleotide primers 5′-ACTCATCACGTCATCACCAAT-3′ and 5′-CTCTCTGCAGCCCTCATTTG-3′ resulting in distinct fragment sizes due to the presence or absence of the deletion/insertion polymorphism.Genomic DNA was extracted from peripheral blood leukocytes according to established protocols. Genotyping of SNPs was achieved by direct sequencing, restriction enzyme digestion of PCR products, TaqMan SNP Genotyping  or primer extension of multiplex PCR products with detection of the allele-specific extension products by the matrix-assisted laser desorption/ionization time of flight (MALDI-TOF) mass spectrometry method  . Logistic regression analyses were carried out using the R software i the genotype of the ith SNP. Here, “a” denotes a constant that centers the risk score distribution around zero and bi relates to the ith variant. The odds ratio of the effect of the ith variant is thus given by exp(bi) Based on the intercept “a” and the single-SNP beta-coefficients estimated using the logistic regression model including all SNPs at once, the genetic risk score (GRS) was calculated asTo estimate the ability of a potential genetic screening test to discriminate between AMD cases and healthy subjects, we computed the receiver-operating-characteristic (ROC) curve. This involves ranking all subjects according to their GRS starting with the smallest, computing sensitivity and specificity at each possible GRS cut-off, and plotting sensitivity versus 1-specificity. The area-under-the-curve (AUC) is a measure of how well the GRS cut-offs can separate AMD cases from controls. We used the package EPICALC rd of the data to the model building (to compute the effect sizes and thus establish the GRS model) and 1/3rd of the data to testing   . We repeIn order to derive the fraction of cases in the five GRS categories as expected in the general population (corresponding to the absolute AMD risk) from the number of cases (N_cases = 986) and controls (N_controls = 796) in our case-control study, we weighted the number of AMD cases in our study byFigure S1Risk estimates for 16 AMD associated variants by disease subtypes. Logistic regression models were fitted with all patients (N = 986), GA cases only (N = 229), NV cases only (N = 581) or mixed GA+NV cases (N = 176) versus controls (N = 796). Odds ratio estimates (OR) are given per risk allele; horizontal bars indicate 95% confidence intervals and the arrow indicates that the boundary extends below 1 or above 6.(TIF)Click here for additional data file.Table S1Published genetic variations associated with AMD.(DOC)Click here for additional data file.Table S2Cross validated absolute risks for late stage AMD in different risk groups in the modeled population.(DOC)Click here for additional data file.Table S3Primers and methods used for genotyping.(DOC)Click here for additional data file.


Vancomycin-resistant enterococci (VRE) are increasing in prevalence at many institutions, especially among patients with co-morbidity conditions that associated with frequent hospitalisation. It is important to determine the risk factors for this resistance microorganism as it will guide the clinician for appropriate antimicrobial therapy and infection control measures.The descriptive, cross sectional study was carried out to determine the prevalence and risk factors for vancomycin resistant enterococci in Kuala Lumpur Hospital (HKL), Malaysia.For 12 months period, antimicrobial susceptibility testing for enterococci species were performed using disk diffusion method. E-test for vancomycin was proceed for the isolates that exhibit resistance to vancomycin by disk diffusion method. To identify the risk factors, a questionnaire was completed for all studied patients and the data were analysed using chi square and multivariate logistic regression.The prevalence of VRE among patients in Kuala Lumpur Hospital was 1%. In chi square analysis, vancomycin usage  showed significance risk factor. In multivariate logistic regression analysis, prolonged hospitalization  and vancomycin use  were associated with potential VRE.The findings of this study will serve as an alert to the clinicians of the emergence of infections by VRE and it will encourage the implemention of appropriate infection control measures and judicious use of vancomycin in order to prevent further rise in VRE prevalence.None declared


A mediation model explores the direct and indirect effects between an independent variable and a dependent variable by including other variables (or mediators). Mediation analysis has recently been used to dissect the direct and indirect effects of genetic variants on complex diseases using case-control studies. However, bias could arise in the estimations of the genetic variant-mediator association because the presence or absence of the mediator in the study samples is not sampled following the principles of case-control study design. In this case, the mediation analysis using data from case-control studies might lead to biased estimates of coefficients and indirect effects. In this article, we investigated a multiple-mediation model involving a three-path mediating effect through two mediators using case-control study data. We propose an approach to correct bias in coefficients and provide accurate estimates of the specific indirect effects. Our approach can also be used when the original case-control study is frequency matched on one of the mediators. We employed bootstrapping to assess the significance of indirect effects. We conducted simulation studies to investigate the performance of the proposed approach, and showed that it provides more accurate estimates of the indirect effects as well as the percent mediated than standard regressions. We then applied this approach to study the mediating effects of both smoking and chronic obstructive pulmonary disease (COPD) on the association between the CHRNA5-A3 gene locus and lung cancer risk using data from a lung cancer case-control study. The results showed that the genetic variant influences lung cancer risk indirectly through all three different pathways. The percent of genetic association mediated was 18.3% through smoking alone, 30.2% through COPD alone, and 20.6% through the path including both smoking and COPD, and the total genetic variant-lung cancer association explained by the two mediators was 69.1%. A mediation model is a statistical approach that explores the direct and indirect effects of an independent variable  on a dependent variable  by including one or more mediating variables (or mediators) Lung cancer GWA studies have consistently shown that the CHRNA5-A3 gene cluster is strongly associated with an increased risk of lung cancer. Also, multiple studies have associated SNPs spanning this region with heavy smoking, nicotine dependence, smoking cessation and COPD The multiple-mediation model used for the study of the SNP, smoking, COPD and lung cancer risk is depicted as a path diagram in To our knowledge, there has been no previous study investigating such a multiple mediation model in the case-control study design setting, in which the standard regression approach could provide biased estimations for the indirect effects as we described above. Therefore, we developed an approach to conduct a multiple-mediation analysis using the model shown in X, 1M, 2M, and Y denote the genetic variant, two mediator phenotypes, and the disease variable, respectively. We assumed binary random variables for both mediator variables and the disease variable, denoted as A and normal allele a. We first assumed a dominant or recessive genetic model for the genetic variant and also denoted it as a binary random variable, a, a) and 1 represents genotypes  and ; for a recessive genetic model, 0 represents genotypes  and  and 1 represents genotype . Note that if an additive genetic model was assumed, a categorical random variable a, a), , and , respectively. Given the random variables, X, 1M, 2,M and Y, the association among all random variables shown in 0a, 0b, 0c, 1a, 2a, 1b, 2b, d, and c′ are regression coefficients and i, j, k = 0, 1. There are different indirect effects in this model 1M, bypassing 2M, which can be assessed as 1b1a (denoted as 1IE); (ii) the indirect effect passing through the mediator 2M, bypassing 1M, which can be assessed as 2b2a (denoted as 2IE); and (iii) the three-path indirect effect passing through both mediators, which can be assessed as 1db2a (denoted as 3IE). Therefore, the total indirect effect passing through the mediators can be given as the summation of the above indirect effects: 1b1a+2b2a+1db2a (denoted as tIE). The regression coefficient c′ represents the effect of the genetic variant on the disease not mediated by either mediator and is usually called the direct effect. In general, the total effect of the genetic variant on the disease is estimated by regressing the disease variable on the genetic variant variable directly. However, the previous analysis has shown that the total effect estimated in this way could be biased when the disease variable and/or mediator variables are binary TE) using an alternative formula defined as the summation of the indirect and direct effects (denoted as TE = tIE+c′). In this case, the percentages of the association explained by the different mediation paths  can be assessed as the specific indirect effects divided by the defined total effect, respectively, and denoted as 1PM = 1IE/TE, 2PM = 2IE/TE, 3PM = 3IE/TE, and tPM = tIE/TE, which represents PM of 1M bypassing 2M, PM of 2M bypassing 1M, PM of both 1M and 2M, and the total PM through different paths, respectively.Let 1a, 2a, and d) will be biased if standard logistic regressions are employed, which in turn, will result in biased estimations of indirect effects and the percent mediated 1a, 2a, and d, we modified the bias-correction approach proposed in our previous study d for the 1M–2M association, while regressing 2M on 1M and X  and 0N are controls (Y = 0) with respect to the disease, the odds ratio (OR) for the association between the mediators 1M and 2M (exp(d)) can be expressed as follows:kjE is the expected number of individuals in the sample, with 2M = k and 1M = j, which is given asj, k, r = 0, 1. The conditional probability kj|rp is written asAs stated above, the regression coefficient 1p and 1q represent the prevalence of the mediator 1M and the disease, respectively, in the general population. The conditional probabilities r|kjp and k|jp are given as functions of regression coefficients:The probabilities 0b, 0c, and d are unknown coefficients of interest. Based on the conditional probabilities given above, we can write the estimated prevalences of the disease and the mediator 2M as follows:N independent individuals for a case-control study of the disease (Y), one can estimate the regression coefficients 1b and 2b as well as the biased coefficient d using logistic regressions based on 0c, 0b, and d. We employed the “fsolve” function in Matlab d for the association between two mediators. As mentioned above, for brevity, the details of correction for the coefficients 1a and 2a were given in 1IE = 1b, 2IE = 2b, and 3IE = 2b.Given a sample with 1 by calculating the OR of SNP random variable X = 1 versus X = 0, which gives the OR for heterozygous genotype against wild-type homozygous genotype. Second, we obtained the biased OR2 by calculating the OR of SNP random variable X = 2 versus X = 0, which gives the OR for homozygous genotype for variant allele against wild-type homozygous genotype. On the basis of OR1 and OR2, and following the different formulas in our previous study When the genetic variant is assumed to be additive, special care needs to be taken. In this situation, we used a categorical random variable, 1M. In this scenario, frequency-matching design also contributes to bias in the estimate of the coefficients for associations among the SNP and the mediators . Therefore, we adapted the approach proposed in our previous work jiE can be calculated asi = 0, 1, 2 and j = 0, 1.Frequency matching is an important and commonly used study design for known risk confounders and has been widely used in case-control studies 1M in the disease cases and controls, given as 1M = 1|Y = 0) prop(1M = 1|Y = 1). In reality, the selection of controls in a frequency-matched study does not have to be perfect, that is, the proportions of individuals with the matched variables do not have to be exactly the same in the disease cases and controls . Given the new formulas for jiE and kjE, one can follow the same procedure described for the unmatched study to assess the corrected coefficients kiE that was used in the unmatched case-control study because the calculation of 1M.If the original disease case-control study is frequency matched on the mediator 1b, and 2b obtained using the standard regression and the corrected coefficients 1IE = 1b, 2IE = 2b, and 3IE = 2b, as well as the total indirect effect tIE = 1b+2b+2b, were obtained by the following steps:Bootstrapping has been employed to evaluate the significance of indirect effects in a multiple-mediator model B samples with replacement from the study data, each with 1n individuals from the disease cases and 0n samples from the disease controls (0+n1n = n). Note that 0n≤0N and 1n≤1N, where 0N and 1N are numbers of cases and controls with respect to the disease in the study sample.Take u = 1, 2, …, B. The corrected coefficients u = 1, 2, …, B are calculated by using the approaches described above.Evaluate the bootstrap regression coefficients using logistic regressions based on the bootstrap samples. Denote the bootstrap coefficients as u = 1, 2, …, B. Let uth ordered bootstrap indirect effects estimations, respectively. Then the 100(1- The bootstrap indirect effects are assessed as Y), the mediator 2M, and the mediator 1M, respectively, which approximate the prevalence values of lung cancer 1a = 0.4055, 2a = 0.4055, d = 0.6931, c′ = 0.4055, 1b = 1.0986, and 2b = 1.0986, which correspond to ORs of 1.5, 1.5, 2, 1.5, 3, and 3, respectively; for the second scenario, we fixed the coefficients as 1a = 0.3365, 2a = 0.3365, d = 0.3365, c′ = 0.6931, 1b = 0.4055, and 2b = 0.4055, which correspond to ORs of 1.4, 1.4, 1.4, 2, 1.5, and 1.5, respectively. The ORs used in this simulation studies were chosen to reflect the observed ORs found in many GWA studies of common human diseases 1M) and different genetic models for the SNP . For the frequency-matched study, we also considered different values for the parameter 1M in disease cases (Y = 1) and controls (Y = 0). On the basis of these parameters, we obtained the values for the intercept regression coefficients 0a, 0b, and 0c for different situations.We performed simulation studies to investigate the performance of our approach for evaluating the indirect effects in the multiple-mediation model in a case-control study . To mimi1M values were then generated on the basis of the dataset of realizations of the SNP using 1M and the SNP values, we used 2M. Last, the disease cases and controls were generated conditional on values of the SNP and both mediators 1M and 2M using Y = 1) and 1,000 disease controls (Y = 0). When a frequency-matched case-control study design with respect to the mediator 1M was considered, the 1,000 disease cases were still sampled randomly. However, the 1,000 controls were sampled so that the proportion of the presence of the mediator 1M in the controls was approximately equal to that in the cases First, we generated genotypes for a SNP using the genotype frequencies, which can be calculated from the MAF. The mediator 1a, 2a, 1b, 2b, c′, and d estimated using both standard logistic regression and the approach proposed in this article are reported in 1M) between the disease cases and controls , the estimated values for c′, 1b, and 2b were 0.4041, 1.0967, and 1.0989, respectively, which were very close to the true values of 0.4055, 1.0986, and 1.0986 used for the simulations. However, the estimated values for 1a, 2a, and d were 0.4615, 0.4547 and 0.7551, respectively, which were biased compared to the true values of 0.4055, 0.4055, and 0.6931. On the other hand, the proposed approach led to estimates of For the unmatched case-control study design, when the standard logistic regressions were applied, the estimates of 1M, in addition to the coefficients 1a, 2a, and d, the coefficient 1b was also highly biased, as expected when the standard regression approach is applied; the coefficients c′ and 2b were still correctly estimated, as in the unmatched study. For example, in scenario one for frequency-matched design, when the proportion of individuals with presence of 1M was higher in cases than in controls by 5% (Δ = −0.05) and the dominant genetic model was assumed, the estimated values of c′ and 2b were 0.4072 and 1.1003, respectively, which were close to the true values of simulation; however, the estimated values of 1a, 2a, d, and 1b were 0.3500, 0.4502, 0.5171, and 0.1189, respectively, which were all highly biased compared to the true values. When we applied the proposed correction approach, however, accurate estimates of When the case-control study was frequency-matched with mediator 1M, bypassing mediator 2M (1IE), the indirect effect through the mediator 2M, bypassing mediator 1M (2IE), the three-path indirect effect through both mediators (3IE), and the total indirect effect, which is the summation of all the specific indirect effects (tIE). We also reported the total effect of the genetic variant on the disease (TE), as well as the percentages of the SNP-disease association explained by different paths . For scenario one, on the basis of the coefficients used for the simulations, the true values of the specific indirect effects and the total effect were given as 1IE = 1b1a = 0.4055×1.0986 = 0.45, 2IE = 2b2a = 0.4055×1.0986 = 0.45, 3IE = 1db2a = 0.4055×0.6931×1.0986 = 0.31, tIE = 1+IE2+IE3IE = 1.20, and TE = tIE+c′ = 1.61, respectively; and the true values of the percent mediated by different indirect effects were given as 1PM = 1IE/TE = 0.45/1.61 = 28%, 2PM = 2IE/TE = 0.45/1.61 = 28%, 3PM = 3IE/TE = 0.31/1.61 = 19%, and tPM = tIE/TE = 1.20/1.61 = 75%, respectively. For scenario two, the true values of the indirect effects, total effect, and corresponding percentages mediated were assessed using the same formulas and were given as follows: 1IE = 0.14, 2IE = 0.14, 3IE = 0.05, tIE = 0.32, and TE = 1.01 for indirect effects and total effect and 1PM = 13%, 2PM = 13%, 3PM = 5%, and tPM = 32% for the percentages mediated through different indirect effects. Similar to 1IE = 0.51, 2IE = 0.50, 3IE = 0.38, tIE = 1.39, and TE = 1.79, respectively, which were all biased compared with the true values of 0.45, 0.45, 0.31, 1.20, and 1.61, respectively. This, in turn, caused the percentages mediated through different indirect effects to be biased as well. For the same example, the percentages mediated by different indirect effects were estimated as 1PM = 28%, 2PM = 28%, 3PM = 21%, and tPM = 77%, respectively. Compared with the true values of 28%, 28%, 19%, and 75%, we can observe that the percentage mediated through the three-way path of both mediators was slightly biased (3PM = 21% versus 19%) and led to a biased estimate for the total percent mediated (tPM = 77% versus 75%). However, by employing the corrected coefficients 1IE = 0.45, 2IE = 0.45, 3IE = 0.31, and tIE = 1.21 for different indirect effects, TE = 1.62 for the total effect, and 1PM = 28%, 2PM = 28%, 3PM = 19%, and tPM = 75% for the percentages mediated through different indirect effects, all of which agreed well with the true values.For the unmatched case-control study, when the standard regression approach was applied, the estimates of the specific indirect effects, as well as the total effect, were biased compared to the true values. This was expected because the coefficients used to assess the indirect effects and total effect were biased. For example, for scenario one with a dominant genetic model (unmatched study), the specific indirect effects and the total effect were given as 1M, the magnitudes of bias in the estimations of indirect effects, total effect, and percent mediated were larger than those for the unmatched study when applying the standard approach. For example, in scenario one for the frequency-matched design, when the proportion of individuals with presence of 1M was higher in the cases than in the controls by 5% (Δ = −0.05) and the genetic model was assumed as dominant, the estimates of indirect effects and total effect were 1IE = 0.04, 2IE = 0.50, 3IE = 0.20, tIE = 0.73, and TE = 1.14, respectively, which were highly biased compared with the true values of 0.45, 0.45, 0.31, 1.20, and 1.61, respectively. Accordingly, the percentages mediated through different indirect effects were estimated as 4%, 43%, 17%, and 64%, respectively, which were also biased compared with the true values of 28%, 28%, 19%, and 75%, respectively. On the other hand, the proposed approach provided estimates of 1IE = 0.44, 2IE = 0.44, 3IE = 0.30, and tIE = 1.18 for different indirect effects, TE = 1.59 for the total effect, and 1PM = 28%, 2PM = 28%, 3PM = 19%, and tPM = 74%, which were all close to the true values.When the case-control study was frequency matched with mediator 1a, 2a, and d in all situations  and biased estimates for the coefficient 1b when the study was frequency matched on the mediator 1M, which in turn led to biased estimates of the indirect effects, total effect, and percent mediated by two mediators. The magnitude of the bias in the estimations increased when the proportion difference . However, the approach proposed in this article can uniformly provide accurate estimates for the coefficients 1a, 2a, and d, and in turn provide accurate estimates for the indirect effects, total effect, and percent mediated for all situations.Therefore, we observed from the overall simulation results that the standard logistic regressions provided biased estimates for the coefficients 1N = 1,153 lung cancer case subjects who were current or former smokers and 0N = 1,137 control subjects frequency matched to the cases by age, sex, and smoking status. All the case and control subjects were Caucasian. Lung cancer cases were accrued at The University of Texas MD Anderson Cancer Center and were histologically confirmed. Controls were ascertained through a multi-specialty physician practice from the same area. Questionnaire data providing information on smoking were obtained by personal interview. This study was approved by the institutional review board at MD Anderson Cancer Center, and all participants provided written informed consent (LAB10-0347). We selected the number of cigarettes per day, or daily smoking quantity (SQ), as the measurement of smoking intensity. The SQ measure is categorized into two levels: SQ<25, light smokers (coded as 0); SQ≥25, heavy smokers (coded as 1) 1M to investigate the mediating effects of smoking and COPD. All the analyses were adjusted for age.We applied our approach to assess the mediating effects of smoking behavior and COPD simultaneously on the association between the SNP rs1051730 and lung cancer risk using a multiple-mediation model  based on2b and c′ should be unbiased, but the estimated coefficients 1a, 2a, and d will be biased. Also, the estimated coefficient 1b for smoking-lung cancer association was statistically non-significant at the 0.05 level of significance  owing to the frequency matching on smoking status. To assess the corrected coefficients i = 0, 1, and 2, were calculated as 0.40, 0.46, and 0.14, respectively. The prevalences of lung cancer . This biased result using the standard approach was not surprising because the case-control data used for this analysis was frequency matched by smoking status . The second indirect effect carries the effect of the SNP on increasing lung cancer risk through only COPD, bypassing smoking behavior. This indirect effect was assessed by the product of 2a and 2b and shown to be statistically significant, as the 95% bootstrap CI did not contain zero . This result means that an individual carrying the deleterious allele for SNP rs1051730 is more likely to develop COPD, and in turn, lung cancer, independent of the individual's smoking behavior. The last indirect effect is the effect of the SNP on lung cancer risk through both smoking and COPD, which is the product of 1a, d, and 2b, and this effect was also statistically significant . This result shows that the individual carrying the deleterious allele is more likely to become a heavy smoker, which in turn causes COPD and then leads to a higher lung cancer risk. The total indirect effect was evaluated as the summation of the three specific indirect effects, and as expected, it was statistically significant . Meanwhile, the direct effect of c′ was also statistically significant , which suggested that the SNP also affects lung cancer risk through a pathway or pathways other than smoking and COPD.When the standard logistic regression approach was applied, not all three specific indirect effects were statistically significant, as evidenced by some bootstrap CIs containing zeros . The firc′) and total indirect (tIE) effects . Given the total effect, we also calculated the percentage of the SNP-lung cancer association explained by each of the specific indirect effects. The percentages mediated were estimated as 3.7%, 37.5%, and 25.1% for the three specific indirect effects, respectively, suggesting that the path through smoking alone explains 3.7% of the SNP-lung cancer association, the path through COPD alone explains about 37.5% of the association, and the path through both smoking and COPD explains about 25.1% of the association. Thus, the results obtained from the standard approach suggested that the SNP influences the lung cancer risk indirectly through two pathways (COPD only and both smoking and COPD) but not through the smoking only pathway. However, this conclusion is likely to be biased, as our simulation results showed that the case-control study design could introduce bias in the estimations of indirect effects, and furthermore, frequency matching on the basis of smoking status could conceal the true underlying association.The total effect of the SNP on lung cancer risk was calculated as the sum of the direct ) and was found to be equal to 1IE = 0.1385, 95% CI = 0.0601, 0.2195, which was statistically significant. This result suggested that an individual carrying the deleterious allele is more likely to become a heavy smoker and, in turn, have a higher risk of lung cancer, independent of the individual's COPD risk. The other two indirect effects were calculated as the products of 2b and 2b, respectively, and were also statistically significant . The total indirect effect  and the total effect  were higher than those obtained from the standard approach, mainly due to the significant indirect effect through smoking only. The percentages of the SNP-lung cancer association explained by each of the specific indirect effects were also calculated and suggested that the path through smoking alone explains about 18.3% of the association, the path through COPD alone explains about 30.2% of the association, and the path through both explains about 20.6% of the association. The results obtained from the proposed approach showed that the SNP rs1051730 influences lung cancer risk indirectly through all three pathways and suggested that a higher percentage of the SNP-lung cancer association was explained by the two mediators with three pathways than indicated by the results obtained from the standard regression approach (69.1% versus 66.3%).Therefore, we applied the new approach proposed in this article to estimate the indirect effects of smoking and COPD on the association between the SNP and lung cancer risk see . The indIn this study, we investigated the multiple-mediation model involving a three-path mediating effect using data from a case-control study. Such multiple-mediation models have been studied previously but not in the context when the study subjects are sampled according to case-control design We applied the approach to investigate the mediating effects of smoking and COPD on the association between the SNP rs1051730 and lung cancer risk using lung cancer case-control GWA study data where the multiple-mediation model was employed. We concluded on the basis of the results obtained from the proposed approach that the SNP rs1051730 influences lung cancer risk indirectly through all three pathways: through smoking only, bypassing COPD (18.3%); through COPD only, bypassing smoking (30.2%); and through both smoking and COPD (20.6%). The percentages mediated through different pathways  obtained using the proposed approach were more correct, according to our simulation results, whereas the percentages mediated obtained using the standard approach were either under-estimated or over-estimated. Our findings that COPD mediates the effect of the SNP on the lung cancer association concurs with a previous study of the association between the SNP rs16969968 (in tight linkage disequilibrium with rs1051730) and COPD One may argue that the use of self-reported, physician-diagnosed emphysema as a COPD measure could result in misclassification of the disease. For example, some studies have shown that when spirometry is used to assess COPD in smokers, estimates of undiagnosed COPD range from 50–80% This study extends our previous work investigating the mediating effects of smoking and COPD on the association between the rs1051730 SNP and lung cancer using a single-mediator model Munafo et al. In our study, we focused on the multiple-mediator model shown in The measure of percent mediated used in our study is usually applicable when the signs of the indirect and direct effects are the same It should be noted that, when we refer to the direct effect, we mean the effect of the SNP on lung cancer risk directly or through pathways other than smoking and COPD.In summary, we investigated the multiple-mediator model, which involves a three-way mediating effect from one mediator to another in a case-control study. We proposed an approach to correct the biased estimations of the indirect effects in such models due to case-control study design. The proposed approach can provide accurate estimations for indirect effects and percent mediated. It is also robust to the case-control study being frequency matched on one of the mediators. The application of the proposed multiple-mediation approach to the study of the association between SNP rs1051730 and lung cancer risk suggests that the SNP has an indirect association with lung cancer risk mainly through its effect on both smoking behavior and COPD, as well as a relatively weaker direct association with lung cancer risk. Currently, several studies are ongoing to identify genetic variants associated with smoking behaviors and COPD using existing GWA study data collected for lung cancer using simplistic regression analyses. Such studies should use more sophisticated statistical models that take into account the complex interplay of smoking, COPD, and lung cancer. Finally, additional studies that include metabolomics markers, and biochemical assays of lung carcinogens as suggested by Spitz et al. Text S11a and 2a for unmatched case-control study design.Correction of coefficients (DOCX)Click here for additional data file.


Acute pancreatitis is the most common complication of diagnostic and therapeutic endoscopic retrograde cholangiopancreatography (ERCP). Several clinical trials used glyceryl trinitrate (GTN) to prevent the incidence of post-ERCP pancreatitis (PEP). However, the results were still controversial.To conduct a meta-analysis of published, full-length, randomized controlled trials evaluating the effect of prophylactic GTN on the prevention of PEP, improve the rate of cannulation and the prevention of hyperamylasemia.Literature searches were conducted using PubMed, EMBASE, The Cochrane Library and Web of Knowledge databases, using keywords "post-ERCP" and "pancreatitis" and limited in randomized controlled trials.Twelve RCTs involving 2649 patients were included. Eleven RCTs compared GTN with placebo for PEP prevention. Meta-analysis showed the overall incidence of PEP was significantly reduced by GTN treatment . Nevertheless, GTN administration did not decrease the incidence of moderate to severe PEP . Subgroup analyses revealed that GTN administered by sublingual was more effective than transdermal and topical in reducing the incidence of PEP. Besides, the prophylactic effect of GTN was far more obvious in the group of high PEP incidence than in the group of low PEP incidence. Additionally, the incidence of hyperamylasemia was significantly reduced by GTN treatment . No differences of the successful cannulation rate of bile ducts  attributable to GTN were observed.Prophylactic use of GTN reduced the overall incidence of PEP and hyperamylasemia. However, GTN was not helpful for the severity of PEP and the rate of cannulation. Pancreatitis remains the most common severe complication of endoscopic retrograde cholangiopancreatography (ERCP) . The incThe ideal pharmacological drug should be highly effective in reducing PEP, have a short administration time, well tolerated with a low side-effect profile and cost-effective. Numerous pharmacological drugs of preventing PEP have been investigated, including non steroidal anti-inflammatory drugs (NSAID), diclofenac, ceftazidime, octreotide, protease inhibitors and heparin and so on. However, most results were disappointing and currently no pharmacological prophylaxis for PEP is in routine use . In the Literature searches of the electronic databases included PubMed, EMBASE, ISI Web of Knowledge (up to May 2013) and the Cochrane Library . The search terms included “endoscopy,” “ERCP,” “endoscopic retrograde cholangiopancreatography,” “post-ERCP pancreatitis,” “post-endoscopic retrograde cholangiopancreatography pancreatitis,” “pancreatitis,” “PEP,” “cannulation,” “GTN,” “glyceryl trinitrate,” “nitroglycerin,” “glyceryl nitrate,” and “randomized controlled trial”. No language restriction was imposed. The manual searching of reference lists from potentially relevant papers was performed to identify any additional studies that have been missed from using the computer-assisted strategy. The following selection criteria were applied: (1) study design: prospective, randomized, controlled trials; (2) study population: patients undergoing ERCP; (3) intervention: prophylactic administration of GTN; (4) comparison intervention: placebo or no treatment; and (5) outcome measures: the overall incidence of PEP, the incidence of moderate to severe PEP, the successful rate of cannulation and the incidence of hyperamylasemia.Two independent reviewers (D.J.X. and J.X.) assessed the quality score of primary trials according to the Jadad scale . The qua2 statistic. Heterogeneity was considered significant by the Cochran Q test with P<0.05 or by I2 greater than 50% [All statistical analyses were performed using Review Manager . All outcomes were expressed as risk ratio (RR) with 95% CI. Heterogeneity was assessed by visual inspection of a forest plot, the Cochran Q test, and the Ithan 50% ,17. A fi177 potential RCTs were identified through in-depth search. Among them, 41 RCTs were excluded because of duplicate studies and 119 RCTs were excluded based on the titles and abstracts . The remaining 17 were then retrieved for full text review. Finally, twelve fully published RCTs met inclusion criteria and were included in this meta-analysis –14,18–26All trials met at least three criteria for trial quality excepted one trial  Table 3Table 3. heterogeneity =0.39, I2= 5%). So we used the fixed-effects model and found that the administration of GTN was associated with a significant reduction in the overall PEP incidence  . Therefore, we pooled the results by the fixed-effects model. Overall result showed no significant reduction in the incidence of moderate to severe PEP   , showing P=0.16) .2=0%, Pheterogeneity =0.43 and RR 0.78; 95% CI, 0.55-1.10; P=0.16; I2=36%, Pheterogeneity =0.18, respectively), whereas the sublingual route was associated with statistically significantly reduced rates of PEP   . Because2=0%; Pheterogeneity=0.58). However, in trials with a high pancreatitis incidence in control group, there was a significant reduction of PEP in the GTN group . The sensitive analysis by excluding study of unclear PEP definition [heterogeneity =0.38, I2= 7%). We further excluded another trial [heterogeneity =0.42, I2= 2%)  .2=0%, Pheterogeneity =0.77). Altogether, 88.9% (1442/1622) patients had successful cannulation, of which 90% (724/804) was in the GTN group and 87.8% (718/818) in the placebo group . Altogether, 40.2% (121/301) of patients had hyperamylasemia, of which 32.9% (50/152) was in the GTN group and 47.7% (71/149) in the placebo group , we changed to random-effect model. GTN use significantly increased the risk of hypotension . In addition, six studies [heterogeneity = 0.07; I2 = 51%), we used fixed-effect model. GTN use significantly increase the risk of headache . Different routes of GTN administration may have different risk of adverse effect. Based on this hypothesis, we performed subgroup analysis of these trials . In other words, patients who received GTN in the periprocedural period were 33% less likely to have pancreatitis. However, GTN-treated patients did not have a reduction of the development of moderate to severe PEP. Though our findings were similar to three previous meta-analyses published on this topic , our metFour routes of GTN used in the trials: topical, sublingual, transdermal and intravenous. Whether the route of GTN administration affects the clinical efficacy was uncertain. From clinical point of view, three studies ,18,26assThe present meta-analysis suggested a significant benefit of GTN in PEP prevention (P = 0.003). The subsequent sensitive analysis by exclusion of two studies ,26 in whThere were several hypothesis of ERCP-induced pancreatitis but is not completely understood ,28. The NO is a reactive nitrogen species (RNS) while GTN is an NO donor. In pancreas cells, many RNS might reactive oxygen to produce peroxy nitrate and damage the cells. Establishment of redox balance is highly complicated, requiring sophisticated regulation of scavenger bioavailability and of RNS generation. The major cellular RNS scavenger in the pancreas is GSH . The thiol group in the cysteine moiety of GSH accounts for its scavenging power. GSH concentration in the pancreas ranks the fourth highest among the visceral organs. Pancreatic GSH turnover is less only than that in the kidney and liver, which have twofold and fourfold the turnover rates of the pancreas, respectively. Therefore, it appears that the pancreas is “evolutionally prepared” for defense against oxidative stress and removal of RNS. Although GSH is the major cellular antioxidant in the pancreas, other cellular antioxidants are also present in the pancreas. In particular, vitamin C, vitamin E, and vitamin A are present in the pancreas in considerable amounts. These antioxidants may also be responsible for cellular defense against oxidative stress. Therefore, pancreas cells would not be damaged by GTN due to redox balance .There were two ways commonly used to prevent PEP: pharmacological and procedural interventions. ERCP should be avoided in unnecessary or low yield cases, especially when multiple patient-related risk factors for pancreatitis development were found. A number of pharmacological drugs, in particular rectal NSAIDs , have alThe ideal pharmacological drugs should be highly effective in reducing post-ERCP pancreatitis, have a short administration time, well tolerated with a low side-effect profile and cost-effective. Several drugs have shown prospect. However, the vast majority have fallen short of these goals. Therefore, adverse effects of GTN should be weighed against its potential clinical benefit. The main adverse effects were transient hypotension and headache, more frequent in intravenous administration delivery of GTN, but did not severe and responded to conventional treatment. The side effects were significantly more frequent in the GTN group and led to dose reduction or cessation of infusion . TherefoBy summarizing all the relevant RCTs published to date, the present meta-analysis confirmed the efficacy and relative safety of GTN. Patients who underwent ERCP, the sublingual route of administration GTN is recommended and should pay attention to its adverse events.In conclusion, this meta-analysis show that the prophylactic use of GTN have an effective and relative safe intervention for preventing PEP and hyperamylasemia, but show no effect of the severity of PEP and the successful rate of cannulation of bile ducts. Further well-designed placebo-controlled RCTs are needed to confirm the effect of GTN in preventing PEP.Checklist S1PRISMA 2009 Checklist for this study.(DOC)Click here for additional data file.Figure S1Meta-analyses between route of GTN and PEP.Forest plot demonstrated a significant decrease in the incidence of PEP of the sublingual route, but had no significant decrease of the topical or transdermal routes with prophylactic GTN use. CI, confidence interval; M-H, Mantel-Haenszel; GTN, glyceryl trinitrate.(EPS)Click here for additional data file.Figure S2Meta-analyses between GTN and incidence of PEP.Forest plot demonstrated a significant decrease in the incidence of PEP in high incidence of PEP group, but no significant prevention of PEP in low incidence of PEP group with prophylactic GTN use. CI, confidence interval; M-H, Mantel-Haenszel; GTN, glyceryl trinitrate.(EPS)Click here for additional data file.Figure S3Meta-analyses between GTN and hypotension.Forest plot showed significantly increase the risk of hypotension with prophylactic GTN use. CI, confidence interval; M-H, Mantel-Haenszel; GTN, glyceryl trinitrate.(EPS)Click here for additional data file.Figure S4Meta-analyses between GTN and headache.Forest plot showed significantly increase the risk of headache with prophylactic GTN use. CI, confidence interval; M-H, Mantel-Haenszel; GTN, glyceryl trinitrate.(EPS)Click here for additional data file.Table S1Analysis of the side effect profile by route of GTN administration.(DOC)Click here for additional data file.


Chronic plantar heel pain (CPHP) is a generalised term used to describe a range of undifferentiated conditions affecting the plantar heel. Plantar fasciitis is reported as the most common cause and the terms are frequently used interchangeably in the literature. Diagnostic imaging has been used by many researchers and practitioners to investigate the involvement of specific anatomical structures in CPHP. These observations help to explain the underlying pathology of the disorder, and are of benefit in forming an accurate diagnosis and targeted treatment plan. The purpose of this systematic review was to investigate the diagnostic imaging features associated with CPHP, and evaluate study findings by meta-analysis where appropriate.Bibliographic databases including Medline, Embase, CINAHL, SportDiscus and The Cochrane Library were searched electronically on March 25, 2009. Eligible articles were required to report imaging findings in participants with CPHP unrelated to inflammatory arthritis, and to compare these findings with a control group. Methodological quality was evaluated by use of the Quality Index as described by Downs and Black. Meta-analysis of study data was conducted where appropriate.P < 0.001) and that CPHP participants were more likely to have plantar fascia thickness values greater than 4.0 mm . CPHP participants were also more likely to show radiographic evidence of subcalcaneal spur than control participants .Plantar fascia thickness as measured by ultrasonography was the most widely reported imaging feature. Meta-analysis revealed that the plantar fascia of CPHP participants was 2.16 mm thicker than control participants (95% CI = 1.60 to 2.71 mm, This systematic review has identified 23 studies investigating the diagnostic imaging appearance of the plantar fascia and inferior calcaneum in people with CPHP. Analysis of these studies found that people with CPHP are likely to have a thickened plantar fascia with associated fluid collection, and that thickness values >4.0 mm are diagnostic of plantar fasciitis. Additionally, subcalcaneal spur formation is strongly associated with pain beneath the heel.


Patients with knee osteoarthritis patients have problems with walking, and tend to walk slower. An important aim of knee arthroplasty is functional recovery, which should include a post-operative increase in walking speed. Still, there are several problems with measuring walking speed in groups of knee osteoarthritis patients. Nevertheless, test-retest reliability of walking speed measurements is high, and when the same investigators monitor the same subjects, it should be possible to assess the walking speed effects of knee arthroplasty. The present study reports a meta-analysis of these effects.A total of 16 independent pre-post arthroplasty comparisons of walking speed were identified through MEDLINE, Web of Science, and PEDro, in 12 papers, involving 419 patients.For 0.5–5 months post-operatively, heterogeneity was too large to obtain a valid estimate of the overall effect-size. For 6–12 and 13–60 months post-operatively, heterogeneity was absent, low, or moderate (depending on estimated pre-post correlations). During these periods, subjects walked on average 0.8 standard-deviations faster than pre-operatively, which is a large effect. Meta-regression analysis revealed significant effects of time and time squared, suggesting initial improvement followed by decline.This meta-analysis revealed a large effect of arthroplasty on walking speed 6–60 months post-operatively. For the first 0.5–5 months, heterogeneity of effect-sizes precluded a valid estimate of short-term effects. Hence, patients may expect a considerable improvement of their walking speed, which, however, may take several months to occur. Meta-regression analysis suggested a small decline from 13 months post-operatively onwards. Patients with knee osteoarthritis have problems walking, and tend to walk slower than controls. Functional recovery is an important aim of unicompartmental (UKA) or total (TKA) knee arthroplasty in patients with symptomatic osteoarthritis, and walking speed may be a useful variable for assessing the functional effects of knee arthroplasty.Over the last years, walking speed has received considerable attention in the literature. In elderly subjects, a decrease in comfortable walking speed may be a sign of co-morbidity , or evenUnfortunately, there are problems in measuring walking speed in groups of knee osteoarthritis patients. Questionnaires are often used, but may be insufficiently valid, since post-operative patients tend to overestimate their own performance when pain has decreased ,12. CleaThere is a vast amount of literature on prognostic factors in knee arthroplasty. For instance, co-morbidity  and highTest-retest reliability of walking speed is high, with most reported IntraClass Correlations (ICCs) at or above 0.9 ,22. ThusThe present study is a meta-analysis, including a meta-regression analysis, of the effects of knee arthroplasty on walking speed. We hypothesized a) large variance in the first period after arthroplasty (due to variability in post-operative recovery), but b) still a clear effect, which, however, c) would decrease after some time (due to co-morbid conditions or an increase in age-related diseases).In August 2009, a search was conducted in MEDLINE, Web of Science, the Cochrane Library, and PEDro, with combinations of the search terms: knee, osteoarthritis, walking, gait, velocity, speed, replacement, arthroplasty, and surgery. Full English reports were included of studies on knee osteoarthritis patients who underwent knee arthroplasty, with mean values and standard-deviations of pre- and post-operative walking speed. To decide on relevance, two members of our research group (H.A.-B. and H.R.F.-Y.) inspected titles and abstracts of all papers, and selected “eligible” studies , with reTwo authors (H.A.-B. and O.G.M.) extracted all relevant data from the papers selected. The number of subjects was registered, plus potentially relevant variables—UKA vs. TKA, distance walked, gender , age, BMI, and any measures of disease severity, co-morbidity, pain, or function. For all studies, self-selected (comfortable) walking speed (mean ± SD) before and after arthroplasty was entered. For descriptive purposes, absolute numbers were used. For walking speed, weighted means were calculated, with standard error of the mean as weighting factor. Since standard-deviations were not always given for age and BMI, these variables were weighted with the number of subjects per study.ESs) were calculated as [Effect-sizes . Open dindependent comparisons, labeled as “studies” .t-value of a paired t-test, corresponding to an rP of 0.537 [None of the papers provided a correlation between pre- and post-operative values. One paper reported , the t-vof 0.537 , which wEffect-sizes were calculated for all comparisons, and depicted against post-operative measurement time Figure . The pat4 = 16 possible meta-analyses. These were all performed. Heterogeneity was always significant , and random models were used. The mean ES equaled 0.21 (range −0.03 to 0.45), with P-values from 0.87 to < 0.001. In 4 of the analyses, the effect-size was significant (P < 0.05). Since between-study variance was large, with mean I2 85.1% (84.8–85.7%), no valid estimate of the short-term effect-size could be reached.Ten studies reported short-term walking speed. Effects varied from significantly slower to significantly faster Figure . In fourQ-values < 6.0, P-values > 0.7), and fixed effect models were used. Both estimates of the middle-term effect-size were significant . In the studies used for these analyses, walking speed increased from 0.96 m/s pre-operatively to 1.16 m/s after 6–12 months. Since I2 equaled 0%, there was no sign of between-study heterogeneity.Ten studies gave mid-term walking speed. Effects varied from non-significantly to significantly faster Figure . In one Q = 3.2, P > 0.1), and a fixed model was used, which revealed the effect-size to be statistically significant (P < 0.001). In the studies used for this analysis, walking speed increased from 0.80 m/s pre-operatively to 1.02 m/s more than 12 months post-operatively. I2 equaled 5.5%, implying low heterogeneity [The four studies with measurements more than 12 months post-operatively were independent. Effects varied from non-significantly to significantly faster Figure . The lonogeneity .rP-values 0.0–0.9 provided largely similar results. For the first five months post-operatively, heterogeneity was sometimes lower (70%–75% for rP 0.0–0.1), but still mostly “high”. For later measurement times, ESs differed by not more than 7% from the initial estimates. The original calculations were used for the conclusion. For 6–12 months post-operatively, heterogeneity was sometimes non-zero (3%–28% for rP 0.8–0.9), but still “low”. There was, however, one noticeable difference, for 13–60 months post-operatively, with heterogeneity turning from “low” to “moderate” (41%–54%) for rP 0.8–0.9.Rerunning the above analyses for B = 0.39, P < 0.001), with larger effect-sizes at later times. Significant positive regression with time was also found in the reruns with correlations 0.0–0.9.In the first five months post-operatively, between-study heterogeneity was high, and we performed univariate meta-regressions of post-operative time (months), age (years), distance walked (m), UKA vs. TKA (1 vs. 0), gender , BMI, and pre-operative walking speed (m/s). We used the first post-operative measurements of the relevant studies. In the initial calculations Table , the onlFor 6–12 months post-operatively, heterogeneity remained low upon recalculation. Therefore, we did not calculate a meta-regression for this time period.B = 0.06, P = 0.02) and a negative effect of time squared , suggesting an initial increase of the effect-size, but a small decrease later on. In all reruns, this pattern remained similar.For measurements 13–60 months post-operatively, heterogeneity was low or moderate. This may warrant meta-regression, but  there weWe performed meta-analyses of 16 studies of self-selected walking speed in 419 patients with knee osteoarthritis before versus 0.5–5 (short-term), 6–12 (mid-term), and 13–60 (long-term) months after unicompartmental or total knee arthroplasty. None of the studies used an untreated control group.P < 0.001).In the short term, between-study heterogeneity was too large for a valid estimate of the overall effect-size. Mid-term heterogeneity was absent or low, and long-term was either low or moderate. In both latter measurement periods, patients walked 0.8 standard-deviations faster than pre-operatively (P < 0.001). Moreover, meta-regression of the last measurement points of all studies revealed a positive effect of time, and a negative effect of time squared , suggesting an initial increase over time, and a decrease later on.In meta-regression of the short-term data, later measurement time coincided with more effect , and a negative effect of time squared . Hence, the present paper confirmed the existence of functional decline after a mid-term plateau [As predicted, meta-regression over the last measurement points of all studies suggested a long-term decline of walking speed. Since the last measurement point may have had a large influence on this result Figure , we reca plateau . On the  plateau . TherefoThe present study is, to the best of our knowledge, the first meta-analysis of the effects of knee arthroplasty on walking speed.Patients tend to have overoptimistic expectations of functional recovery after arthroplasty, and to underestimate recovery time . MoreoveThe physiology of comfortable walking speed remains largely unknown. Just as in breathing rate, it is easy to change walking speed at will, but, again like breathing rate, it is prone to fall back to its own intrinsic parameters. Walking speed is a fair predictor of, e.g., co-morbidity, atherosclerosis, inflammatory status, cognitive impairment, hospitalization, and even mortality , and matTraditionally, UKA was used for older, inactive patients with medial knee osteoarthritis . To dateFour databases were used  with a limited number of search terms. We may have missed relevant papers, authors who found no effects on walking speed may have refrained from reporting it, and other forms of publication bias cannot be excluded. Note, moreover, that the two authors who read the papers in detail did not do so independently. On the other hand, this was a relatively modest study, with straightforward results, that were in agreement with the clinical literature.Q-statistic in the choice for a random or a fixed effects model [Q, and I2, all led to the same classification of heterogeneous vs. non-heterogeneous subsets of effect-sizes, suggesting that the specific calculation that was used did not affect main results.There is still some debate in the literature on the appropriateness of using the standard ts model . In the An important limitation of the present study is the lack of randomized controlled trials, which precluded causal interpretation. Correlations between pre- and post-test were never explicitly provided, but recalculations suggested that this omission did not affect the pattern of results. The number of studies analyzed was relatively low, and potentially relevant factors were often not mentioned, which made it impossible to reach any firm conclusions as to which factors enhance rapid or lasting functional recovery.A meta-analysis of pre and post-arthroplasty walking speed revealed a large effect 6–60 months post-operatively. For the first 0.5–5 months post-operatively, heterogeneity of effect-sizes precluded a valid estimate of short-term effects. Hence, patients may expect a considerable improvement of their walking speed, which, however, may take several months to occur. Moreover, the analysis suggested a small decline from 13 months post-operatively onwards. Such a decline may be a sign of increasing co-morbidity.The authors declare that they have no competing interests.All authors made substantial contributions to conception and design of the study. HAB, HRFY, and OGM were involved in data acquisition, HAB, HRFY, OGM, HCWdV, SMB, DLK, and JHvD were involved in data analysis and interpretation of data. All authors were involved in drafting the manuscript, and revising it critically. All authors have given final approval of the version to be published.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2474/13/66/prepub


RAS mutation (RASmut) in acute myeloid leukaemia (AML), which has been associated with intra-uterine chemical exposures. A case-case study was conducted in order to explore maternal and child exposures to tobacco smoking associations with early age leukaemia (EAL).Deregulation of the MAPK genes signalling caused by somatic mutations have been implied in leukaemia pathogenesis, including MLL rearrangements (MLL-r), RASmut and NQO1 rs1800566 (C609T). Samples from 150 acute lymphoblastic leukaemia (ALL) and 85 AML were included. Maternal exposures were assessed using a structured questionnaire with demographic, personal habits and residence history information. Restriction fragment length polymorphism and denaturing high performance liquid chromatography were used to screen FLT3, KRAS, and NRAS mutations; direct sequencing was performed to validate the results. NQO1 polymorphism was detected by real-time allelic discrimination technique.Covariables of reference were RASmut were detected in 28.7% of EAL cases; BRAFmut was found only in one AML patient. Higher rate of KRASmut was found in ALL (30.3%) compared to AML (20.8%) with MLL-r; RASmut showed an association with second-hand tobacco smoking exposures . A considerable increased risk for EAL with the combination of RASmut and NQO1 609CT  was observed.Overall, MLL-r. Additionally, suggests that children second-hand tobacco exposures are associated with increased risk of EAL with RASmut modulated by NQO1 rs1800566 (C609T).Our data demonstrated the increased risk association between maternal smoking and EAL with  MLL rearrangements (MLL-r) and ETV6-RUNX1 fusion genes  that focuses on the investigation of EAL ,23. Part. The NQO1 status was considered as at least one T allele variant for analysis [The design is a case-case study in which ALL were compared with AML according to distribution of gene mutations, age at onset of the disease, parental’s demographics and maternal exposures to tobacco smoking. The Pearson’s chi-square and Fisher exact tests and child profile were used to assess differences in discrete variables. Crude and adjusted odds ratio (OR) and their 95% confidence intervals (CI) for selected variables  were assessed using unconditional logistic regression in order to estimate the magnitude of associations. Association tests between biological variables and parental smoking habits before, during and after pregnancy were performedanalysis .RAS mutation, NOQ1609CT, presence of MLL-r and presence of smoker within the household (SMOKER) and two-by-two interaction terms between them. Full log-linear model: Log(count) = a + b1(Var1) + … + b4(Var4) + c1(Var1 * Var2) + c2(Var1 * Var3) + c3(Var1 * Var4) + d1(Var2 * Var3) + d2(Var2 * Var4) + e1(Var3 * Var4). A sub model without each of the interaction terms was compared to the full model containing all the pairwise interactions by performing the log-likelihood test and generating a P- value, allowing testing if the pairwise associations between the variables was relevant in the study population. The log-linear analysis is an extension of the Pearson’s χ2 test, consisting of and additive part of each variable main effects combined with the pairwise multiplicative interaction terms. For all analyses, the p values <0.05 were considered statistically significant.Following previous studies exploring the association of different variables with the occurrence of paediatric leukaemia , a multiRASmut and RAS wild-type (RASwt) in the children outcome. Patients lost to follow-up were censored at their date of last known contact. The statistical Software SPSS Version 18.0 was used .An estimate of overall survival (OS) was determined using the Kaplan–Meier and log rank tests in order to verify the association of The demography and clinical characteristics of EAL are shown in Table MLL-r could be detected; the MLL status indeterminate category includes those with insufficient biological material ; the remaining 109 cases had the MLL status defined as negative. Somatic chromosomal aberrations, such as hyperdiploidy , translocation with the fusion transcript ETV6-RUNX1  TCF3-PBX1 , t and t and complex karyotype  were found in ALL cases older than 12 months of age at the diagnosis. RUNX1-RUNX1T1 and CBFβ-MYH11 fusion transcripts were found in 6 AML cases; one child with AML-M3 presented PML-RARa fusion transcript.In 102 out of 138 cases (73.9%) the presence of an RAS mutations were found in 60 (25.5%) of the acute leukaemia cases, whereas FLT3 mutations were found in 11 patients (6.9%) predominantly among ALL and not statistically significant. One hundred twenty-five cases were successfully analysed for BRAF mutations and only one AML case harboured a mutation located at c.1799 T > A (V600E). Concurrent mutations in FLT3 and RAS in the same patient were detected in six patients; two of these cases harboured an MLL-AFF1 fusion gene. The demographic characteristics and the distributions of KRAS, NRAS, and FLT3 mutations were not statistically different according to age strata  represented 20.0% and NRAS  16.1%. While FLT3 mutations  were equally distributed in infants and children older than 12 months, KRASmut was strongly associated with age less than 12 months and MLL-r (p = 0.001), as shown in Additional file nd in 60 5.5% of tMLL-r and the frequencies of RAS and FLT3 mutations in EAL  in ALL and in AML  were not significant to be included in further analysis. Only one case presented a BRAF mutation, demonstrating that is a rare alteration in EAL.We have tested acute leukaemia subtypes associated with RASmut and RASwt among ALL and AML according to patient age at diagnosis  in ALL , shown in Additional file RASmut and MLL-r showed a similar poorer outcome; high OS for RASmut cases presenting MLL wild type (MLLwt) was also found  allele frequencies in early age ALL and AML are shown in Table RAS mutations and NQ01 rs1800566 (C609T) genotype are presented in Additional file NQO1T609T and RAS mutation  without statically significance. The association of NQO1 609CT, RAS and MLL status in these settings of childhood leukaemia were further tested and results are presented in Additional file NQO1 rs1800566 (C609T) variant presented an increased risk association with ALL with MLL-r, although not statistically significant (p = 0.09).The distributions of the Information on the lifelong length of tobacco exposure, smoking during pregnancy and second-hand smoking were obtained and reported previously; no dose–response effect was found . The cohRAS status, NQO1 polymorphism, and EAL with MLL-r are shown in Table MLL-r .The risk association of tobacco smoking exposure during pregnancy, with the RAS status and NQO1 polymorphism. However, an increased risk association was observed with NQO1609CT and the presence of someone smoking at home during the EAL pregnancy .Moreover, 54 (45.5%) mothers reported the presence of someone smoking at home during the index pregnancy (second-hand smoking). No risk associations with mothers who ever smoked (before and/or during pregnancy) were observed with NQO1 and RAS status was tested ; the association of second-hand smoking during pregnancy and children with NQO1 rs1800566 (C609T) polymorphism increased 2.97 folds the risk of developing leukaemia ; RASmut and the presence of at least one variant allele of NQO1 rs1800566 (C609T) showed an increased risk developing leukaemia . No associations were observed with FLT3 mutations and maternal exposures (data not shown). The results of the multivariable log-linear model and to measure the association with selective variables shown in Table RASmut and second-hand smoking and the NQO1 rs1800566 (C609T) polymorphism and second-hand smoking as EAL risk factors.The risk association of smoking exposures during pregnancy combining ed Table . The preChildhood leukaemia is resultant of a multiple causation factors. The exogenous or endogenous exposures, the intrinsic susceptibility of the child given by single-nucleotide polymorphism may increase the chances of childhood leukaemia ,8. This RASmut 25.5%) in these EAL settings associated with MLL-r. The prognostic value RASmut in childhood leukaemia with or without MLL-r was described and demonstrated the poor risk associations [First of all, the analyses took into account the relatively high prevalence of MAPK mutations  that some mothers might had denied being a smokers due to guilt. This is one pitfall of epidemiological studies based on questionnaire responses. Censured topics and variables such drug and/or tobacco users values are depending upon the commitment of different actors under social pressures. However, the biologic plausibility for the present findings is that cigar metabolites compounds carcinogens’ substances such phenols, formaldehyde, toluene and others co-carcinogens (catechol and related compounds), toxic agents  and free radical species (nitric oxide and others) that are spread in the air by the smokers are contaminants with DNA damage power. The carcinogenesis pathways of these substances have already been determined . MetayerNQO1 gene involved in benzene activation and participating in benzoquinone detoxification and reactivating benzene intermediates, might contribute to outgrowth extrinsic effect in exposed cells [NQO1 rs1800566 heterozygous individuals (C/T) have intermediate enzyme activity, and homozygotes for the variant allele (T/T) are deficient in NQO1 activity [MLL/AFF1 fusion genes [NQO1 rs1800566 (C609T) according to age strata in EAL and control groups in Brazilian settings were tested and the genetic susceptibility data regarding NQO1 variants are described elsewhere [NQO1 C609T was detected in children older than 1 year, although this association lacked statistical significance , whereas, a protective effect of the NQO1 C609T variant was found for infants with MLL germ line acute leukaemia  [Molecular epidemiology suggests that the risk conferred by gene variants have decreased the expression of specific products, but contribute to the intrinsic vulnerability of immature cells to initiate an abnormal clone. In this context, the polymorphism of ed cells . NQO1 rsactivity . Accordion genes -44. The lsewhere . In brie16-0.81) .RASmut and NQO1 variants showed a crescent association with the presence of someone smoking in the house suggests the synergistic effect of both genetic alterations contributing for leukaemogenesis. At such age strata, with a short postnatal life span of exposures, the importance of exploring gene mutations and environmental exposures is of paramount importance, because it allows to speculate that despite of the timing of tobacco smoked during pregnancy the dose-effect was modulated by genetic susceptibility with the potential of causing EAL. These results add evidence to our hypothesis that RASmut might be consequence of hazardous exposures in individual with genetic susceptibility.The observation that the combination of KRASmut was observed to be strongly associated.The current investigation has some limitations, as consequence of case-case study in such rare disease. First, the reduced numbers of children with low frequency of maternal exposures make odds ratios unstable and may have incurred in insufficient statistical power to detect significant differences. Second, the selection of cases due to biological material available may have introduced selection bias. However, the comparison of the analysis of the biological samples from children with and without maternal interview was similar. Another weakness of our study that should be pointed out is the missing assess of parental occupational exposures that involve chemical substances such as dyes, polycyclic aromatic hydrocarbons, organic powder/industrial dust, in which, On the other hand, the study presents strong points. At first, it allowed the collection of maternal data exposure to smoking in a very rare disease setting such as infant leukaemia and combines the results with somatic gene mutations and inherited predisposition via gene variant related to smoking exposure.RAS and NQO1 polymorphism in EAL with MLL-r.Finally, to our knowledge, this is the first study exploring the association between maternal exposure to tobacco smoking during pregnancy and MLL-r. Additionally, suggests that second-hand tobacco smoking exposures are associated with increased risk of EAL with RASmut modulated by NQO1 rs1800566 (C609T). The significant associations found here could guide the design of other etiological studies in childhood leukaemia, emphasizing the critical role of genetic susceptibility with somatic mutations in the mechanistic pathway leading to leukaemia in childhood.The present data demonstrated the increased risk association between maternal smoking and EAL with MLL-r: MLL rearrangements; MLLwt: MLL wild type; NQO1: Detoxification enzyme NAD(P)H: quinone oxidoreductase; OR: Odds ratio; OS: Overall survival; PSQ: Pyrosequencing; RASmut: RAS mutation; RASwt: RAS wild-type; RFLP: Restriction fragment length polymorphism; RT-PCR: Reverse transcriptase polymerase chain reaction; WBC: White blood cell.ALL: Acute lymphoblastic leukaemia; AML: Acute myeloid leukaemia; CI: Confidence intervals; dHPLC: Denaturing high-pressure liquid chromatography; EAL: Early age leukaemia; FISH: Fluorescence in situ hybridisation; ITD: Internal tandem duplication; MAPK: Mitogen-activated protein kinase; The authors declare that they have no competing interests.FGA, JMFS, BAAG and ME analyzed and interpreted the data and wrote the manuscript. FGA, JMFS, BAAG and TCB performed the laboratory work of this study. FGA, JMFS and LCST participated in the statistical analysis. MSPO contributed to the conception and the design of the study, revised the article critically, and made the final approval of the version to be submitted. BCSGIAL contributed with samples. All authors’ read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2407/14/133/prepubDemographic and clinical distribution of variables analysed in childhood acute leukaemia according to maternal enquires, Brazil, 2000-2010.Click here for fileDemography and laboratorial characteristics of earl age leukaemia according to age groups, Brazil.Click here for fileRAS mutations, according to age strata, Brazil, 2000-2010.The demography and laboratorial differences in the distribution of Click here for fileRAS and FLT3 mutations in early age leukaemia according to MLL status, Brazil 2000-2010.The association of Click here for fileRAS mutations (K- or N-RAS) and RAS (K- or N-RAS) wild-type among ALL and AML according to patients age at the diagnosis. (A) RAS mutations in ALL cases aged ≤12 months. (B) RAS mutations and MLL rearrangements in ALL cases aged ≤12 months. (C) RAS mutations in ALL cases aged 13–24 months. (D) RAS mutations and MLL rearrangements in ALL cases aged 13–24 months. (E) RAS mutations in AML cases aged ≤24 months. (F) RAS mutations and MLL rearrangements in AML cases aged ≤24 months. ALL: acute lymphoblastic leukaemia; AML: acute myeloid leukaemia.Overall survival of Click here for fileRAS mutations and NQ01 genotype in EAL.The frequency association between Click here for fileNQO1, RAS and MLL status in childhood leukaemia, Brazil 2000-2010.The Click here for file