APOE gene and the onset risk of Parkinson's disease dementia (PDD).  To clarify the relationship between certain genotypes or alleles of the APOE gene and the onset of PDD. Manual information retrieval was also performed. All studies that met the quality requirements were included in a meta-analysis performed using RevMan 5.3 software.  The PubMed, Cochrane, Embase, CBM, CNKI, and Wanfang databases were searched to identify all case-control studies and cohort studies published before October 30, 2017, that investigated the association between the APOE gene on PDD onset was analyzed from three aspects: five genotypes vs. ε3/3, ε2+/ε4+ vs. ε3/3, and ε4+ vs. ε4−. The risk factors for PDD may include the genotypes ε3/4  and ε4/4 . In patients with PDD, there was no significant difference in the distribution of ε2+ vs. ε3/3 . The risk of PDD was 1.61 times greater in ε4+ compared with ε3/3 . As the results indicated that ε2+ did not play a role as a risk factor or a protective factor, we divided the population into ε4+ and ε4− for the meta-analysis and found that, among patients with Parkinson's disease, the dementia risk of those with ε4+ was 1.72 times greater than that of those with ε4− . Subgroup analysis in accordance with different geographical regions revealed that ε4+ was a risk factor for PDD in people from all regions.  The meta-analysis included 17 studies, with a total of 820 patients in the PDD group and 1,922 in the non-PDD group. The influence of the APOE genotypes, ε2+ is neither a risk factor nor a protective factor for PDD, while ε4+ is a risk factor for PDD. The present results are applicable to Asian, European, and American patients with Parkinson's disease. Regarding the single APOE genotypes, ε3/4 and ε4/4 may be risk factors for PDD; however, further studies with large sample sizes are needed to verify this. Among the  Parkinson's disease (PD) is a common neurodegenerative disease among middle-aged and older adults. The major clinical features of PD include motor symptoms , as well as nonmotor symptoms . Parkinson's cognitive disorders are a common nonmotor symptom of PD, and these can be divided into mild cognitive impairment and Parkinson's disease dementia (PDD). An epidemiological investigation performed in 2005 showed that dementia develops in 24–31% of patients with PD and that PDD accounts for 3-4% of patients with all types of dementia [APOE, MAPT, SNCA, GBA, LRRK2, and COMT genes have been found to play a role in the onset and development of PDD [The clinical features of PDD include insidious onset and slowly developed deficits of attention, executive function, visual spatial function, and memory, accompanied by illusion, delusion, indifference, and other spiritual and emotional changes . The patt of PDD .APOE gene than any other gene, and the APOE gene is generally regarded as the gene that has the largest influence on dementia and a stronger predictability compared with other genes.The risk factors for the onset of PDD are likely to be related with the presence of specific genes, and the presence of a certain genotype or an allele may predict whether the risk of PDD is increased in patients with PD. This would enable the risk factors for the onset of PDD to be predicted through testing for related genes, and people identified as being at high risk of PDD could promptly commence tracking and prevention therapies to prevent PDD and suspend its progress. Regarding the research into various genes related to the risk factors for the onset of PDD, a greater number of studies have evaluated the APOE gene has ε2, ε3, and ε4 alleles, which can be classified into six different genotypes: ε2/ε2, ε2/ε3, ε2/ε4, ε3/ε3, ε3/ε4, and ε4/ε4. These genotypes can be divided into the E2 phenotype (ε2/2 and ε3/2), E3 phenotype (ε3/3), and E4 phenotype , among which the E3 phenotype is the most common and is referred to as the wild type [APOE) [APOE can influence cholesterol/lipid homeostasis, synaptic function, glycometabolism, neurogenesis, mitochondrial function, tau protein phosphorylation, neuron atrophy, neuroinflammation, and the metabolic and gathering pathways of β-amyloid protein (Aβ) [APOE can also protect the central nervous system by reducing its oxidative stress and inflammatory response level, resulting in cerebral protection [APOE can stimulate neural stem cells to enhance their survival through the conduction path of the extracellular-signal-regulated kinase signal [APOE functions, affecting the abovementioned biochemical reaction processes and causing cognition impairment.The ild type . These tild type , and thee [APOE) . In the ein (Aβ) –16. APOEotection . Furthere signal . DiffereAPOEε2 allele protects the central nervous system, and a longitudinal study found that APOEε2 effectively reduces damage to the parts of the brain that control daily function and episodic memory [APOEε4 causes damage to the central nervous system that can increase the risk of cognitive disorder and is one of the major risk factors for dementia [APOE gene on the risk of onset of PDD. A case-control study investigating the relationships between PDD and different APOE genotypes found no obvious differences in APOE genotypes and gene frequency between patients with versus without PDD [APOEε2+ and ε4+ might carry a higher risk of PDD [APOEε4 increases the risk of PDD and that ε2 has no relationship with dementia development in patients with PD [APOE gene is related to the risk factors for the onset of PDD and to define its risk level.At present, it is widely believed that the c memory . In contdementia . Many sthout PDD . In contk of PDD , while a with PD . The conAPOE gene and the onset of PDD. The aim of the present meta-analysis was to provide a more objective evidence-based medicine foundation for the relationship between different APOE genotypes and the risk factors for the onset of PDD.The present meta-analysis was performed to make a quantitative synthesis and comprehensive assessment of published materials on the association between the APOE gene and the onset of PDD. The references of retrieved articles, conference literature, and gray literature were also searched manually. The search words were Parkinson disease, Parkinson's disease, primary parkinsonism, parkinsonism, primary, paralysis agitans, Parkinson's disease, Parkinson dementia complex, apolipoprotein E, apoprotein E, APOE, APO-E, APO E, AD2, LPG, LDLCQ5, dementia, cognition disorders, cognitive defect, dementias, demention, amentia, amentias, case-control study, and cohort studies.A method combining subject and free terms was applied to comprehensively and systematically search the PubMed, Cochrane, Embase, CBM, CNKI, and Wanfang databases for case-control studies and cohort studies published before October 30, 2017, that investigated the relationship between the APOE gene and the onset of PDD; (2) explicit clinical diagnosis of PD  or pathological diagnosis; (3) APOE genotype recorded; (4) at least one method used to assess dementia; (5) complete description of the results, and the odds ratio (OR) and 95% confidence interval (CI) of the case and control groups could be obtained either directly or indirectly; (6) case-control study or cohort study; (7) published in Chinese or English; (8) full text could be obtained, or the authors could provide the requisite information and data; and (9) published or unpublished materials before October 30, 2017. Exclusion criteria were as follows: (1) failure to match the research aim ; (2) diagnostic criteria of PD were not stated clearly, or unspecialized diagnostic criteria were used; (3) incomplete gene detection records; (4) the method used to assess dementia was not described; (5) abstract, literature review, case report, seminar, or repetitively published literature; for repetitively published literature, the most recent article or the article with most complete data was selected; and (6) full text could not be obtained, or sample data were not complete or clear and requisite information and data could not be acquired after contacting the author.Inclusion criteria were as follows: (1) observational study investigating the relationship between the The Newcastle-Ottawa Scale was used to assess the methodological quality of the included studies . For casAPOE gene, and the exposure assessment method was defined as the method used to detect the gene. Each item that met one of the abovementioned criteria was represented by ∗, and each ∗ was equivalent to 1 point, giving a potential total of 9 points. Higher scores indicated higher quality studies; studies with a score of 6 points or higher were included in the present meta-analysis.Exposure was defined as the allele or genotype of the The following data were extracted and tabulated: first author, publication date, country of the study population, race, age, diagnostic criteria of PD, diagnostic criteria of dementia, study design, sample capacity, and genotype distributions of the case and control groups. The literature screening, quality assessment, and data extraction were completed by two researchers, and disagreements were resolved via discussion with a third researcher.APOE gene and the onset of PDD and to calculate the OR and 95% CI for the analyses of the five genotypes  vs. ε3/3, ε2+/ε4+ vs. ε3/3, and ε4+ vs. ε4−. The Q value and I2 were used to test the heterogeneity. P < 0.10 was considered to indicate heterogeneity between combined studies. I2 values of 0–25% indicated no heterogeneity, 25–50% indicated mild heterogeneity, 50–75% indicated moderate heterogeneity, and 75–100% indicated major heterogeneity [Z was used to test and calculate the significance of the OR value and was the criterion used to evaluate risk correlation. The applied inspection level was a = 0.05, and P < 0.05 was considered to indicate a significant difference. Heterogeneity tests were initially done within each group; when heterogeneity was detected, the source of the heterogeneity was investigated via subgroup analysis. The single removal method was applied in the sensitivity analysis to test the stability of the results. Funnel plots were used to test for publication bias.RevMan 5.3 software was used to analyze the relationship between the ogeneity . The staogeneity  was usedogeneity  was usedThere were 426 articles retrieved; screening of the titles and abstracts resulted in the exclusion of 62 articles that were repeated literature, six that were conference literature, and 316 that were incompatible with the research contents of the present analysis. Of the remaining 42 articles, the original text of one article could not be obtained, two articles reported the same experiment (the one with more complete data was included), the experimental grouping in five articles differed from that used in the present meta-analysis, five articles had used undefined or incorrect diagnostic criteria for PD, four articles had no cognitive evaluation criteria, four articles had incomplete or irrelevant data, and the quality assessment score of five articles was less than 6 points. A final total of 17 articles were included in the meta-analysis  and those without the ε4 genotype (ε4−) [APOE gene and PDD [All 17 included studies investigated the relationship between dementia in patients with PD who carried the pe (ε4−) , 30–42.  and PDD , 30–36. APOE gene, the frequencies of some genotypes were low and the event counts were 0, and thus, it was impossible to calculate the OR values separately. Therefore, we calculated OR values and 95% CIs for the five genotypes  vs. ε3/3. As shown in ε3/3 genotype, there was a significantly greater risk of dementia in those with genotypes ε3/4  and ε4/4 (OR 2.93 95% CI 1.20–7.14), while there was no difference in the risk of PDD between those with the ε3/3 genotype and those with the genotypes ε2/2 , ε2/3 , and ε2/4 .In the 10 included studies that investigated all genotypes of the ε2+/ε4+ vs. ε3/3 in 10 included studies are shown in Figures Q testing, P > 0.10), and so the fixed-effect model was selected. The combined OR value in ε4+ patients was 1.61 times greater than that in those with the ε3/3 genotype . ε3/3 genotype had a similar risk of PDD development compared with ε2+ patients .The OR and 95% CI for ε4+ vs. ε4− are shown in I2 = 25%, the fixed-effect model was used. The combined OR value in ε4+ patients was 1.72 times greater than that in ε4− patients , indicating that carrying the ε4 genotype was a significant risk factor for the development of PDD.This part of the meta-analysis included 17 studies, and the OR and 95% CI for APOE gene on the risk of PDD onset, we performed a subgroup analysis in accordance with the regional distributions of patients.To determine whether there were regional differences in the influence of the ε2+ was not a risk factor for PDD development compared with patients carrying the ε3/3 genotype ; for Europeans, the risk of PDD was 1.53 times greater in ε4+ compared with ε3/3 . All subgroups had no heterogeneity , four studying European patients, and one studying American patient. In both Asian and European patients, genotype . Howeverogeneity .ε4 genotype was a risk factor for PDD development, but the degree of risk varied in different regions; the risk of PDD onset in ε4+ patients compared with ε4− patients was increased by a factor of 1.46 in Asian patients , a factor of 1.41 in European patients , and a factor of 2.32 in American patients .In accordance with the geographic distribution of the study populations, the 17 studies were divided into five studying Asian patients (Chinese), six studying European patients, and six studying American patients. ε2+ vs. ε3/3, ε4+ vs. ε3/3, and ε4+ vs. ε4−. After removing the studies included in the analyses of ε4+ vs. ε3/3 and ε4+ vs. ε4−, we performed a meta-analysis of the remaining studies; there were no obvious changes in the combined OR values, and all had statistical significance. Moreover, we did not identify any individual studies that had brought significant heterogeneity into the analysis of various studies. Removing the study published by Wang in 2014 [ε2+ vs. ε3/3 changed the result from having no statistical significance  to having statistical significance ; however, the quality of this study was high, the diagnosis and gene detection methods were standard, the experiment design was reasonable, and the results were reliable [P=0.87, P for heterogeneity = 0.90, I2 = 0%).In Section , we perf in 2014  from theε2+ vs. ε3/3 and ε4+ vs. ε3/3 were basically symmetrical, with all points evenly dispersed on both sides of the central line and basically located within the 95% CI and no unfilled corners. Hence, we considered that the possibility of bias was not large. Inspection of the funnel plot of ε4+ vs. ε4− revealed that the symmetry was good, but that two studies were located outside the 95% CI, indicating that there might be a degree of bias; however, separate removal of these two studies showed that their removal exerted no influence on the result , while it was 1.76 times greater in Asian patients and 1.41 times greater in European patients.The present meta-analysis included 17 studies, comprising 820 patients in the experimental group (PDD group) and 1,922 in the control group (non-PDD group). The PDD onset risks of patients with different genotypes of the APOE genotypes influence dementia development in patients with PD is still unclarified, although many studies have investigated the mechanism by which APOEε4 leads to dementia. APOEε4 participates in the mechanism of dementia via the following four aspects: first, Aβ retention can form age pigmentation and vascular amyloidosis and thus lead to dementia. APOE adjusts the combination of Aβ through lipidation [β in the form of a molecular chaperone to influence the elimination of Aβ. The ability of ε4+ to eliminate Aβ is weaker than that of ε3/3 [APOE gene can participate in the immune adjustment of the central nervous system; the immune response of the central nervous system in ε4+ is stronger than that in ε3+, and excessively strong immune responses can lead to brain injury and dementia [APOE gene subtype can play a regulatory effect in the injury and repair of synapses, and a decline in the number of dendrites in the hippocampus of ε4+ may be related to dementia [The mechanism by which different pidation , and it  of ε3/3 . Second, of ε3/3 . Third, dementia . Fourth,dementia .APOEε4 is one of the risk factors for PDD, and this information can be used to guide the therapeutic direction in patients with PD. Detection of the APOE gene can predict the risk of PDD onset; high-risk patients can then be more closely monitored, intervening measures (such as controlling the risk factors) can be implemented to prevent PDD, and PDD can be diagnosed and treated in the early stage so that the disease progression can be postponed, patients' quality of life can be improved, social and family burdens can be relieved, and the mortality rate can be lowered. It is already known that the onset of Alzheimer's disease (AD) is related to the APOE gene, and the present results indicate that this gene is also a risk factor for the onset of PDD, suggesting that the pathogeneses of PDD and AD may be similar. Although there are only few studies on PDD, we may be able to use the relatively better understood pathogenesis and treatment of AD as a reference to provide ideas for research on PDD. In addition, as the pathogenic factors of AD and PDD may be similar, it is possible that the treatment methods used for AD are applicable to PDD. These theories should be investigated in subsequent studies, for which the results of the present meta-analysis can provide theoretical foundations.The present meta-analysis revealed that Compared with two previous meta-analyses , 48, theAPOE gene and revealed some limitations that may provide future research directions. A reasonable PDD diagnostic method is urgently required, as no studies have investigated the effectiveness of the currently available diagnostic criteria for PDD, so many studies have used unsuitable criteria such as the diagnostic and statistical manual of mental disorders (DSM) and the mini-mental state examination. In addition, future studies should enhance the representativeness and credibility of sample populations, adopt multicenter, multiracial, and larger sample sized community-based control or cohort studies, and further evaluate the link between the APOE gene and the risk of PDD onset among people of different ages and education levels.The present meta-analysis investigated the risk of PDD onset in relation to the presence of the APOE genotypes, ε2+ is neither a risk factor nor a protective factor for PDD onset, while ε4+ is a risk factor for PDD. The present findings are especially applicable to Asian, European, and American patients with PD. Regarding single APOE genotypes, ε3/4 and ε4/4 may be risk factors of PDD, but studies with large sample sizes are needed to verify this.Among the


A detailed examination of the two meta-analyses reveals that the authors employed different approaches, used different inclusion and exclusion criteria, analyzed different sets of studies, described their methods with insufficient detail to compare them clearly, and did not report or properly account for significant small sample size bias. The first objective of the current study was to reanalyze the studies selected in each of the published meta-analyses, while taking into account small sample size bias. The second objective was to replicate each meta-analysis by extracting relevant effect sizes directly from the primary studies included in the meta-analyses. The present study revealed three key findings: (1) many of the primary studies used a small sample size; (2) small sample size bias was pronounced in many of the analyses; and (3) when small sample size bias was taken into account, the effect of PPIs on well-being were small but significant (approximately r = .10), whereas the effect of PPIs on depression were variable, dependent on outliers, and generally not statistically significant. Future PPI research needs to focus on increasing sample sizes. A future meta-analyses of this research needs to assess cumulative effects from a comprehensive collection of primary studies while being mindful of issues such as small sample size bias.For at least four decades, researchers have studied the effectiveness of interventions designed to increase well-being. These interventions have become known as positive psychology interventions (PPIs). Two highly cited meta-analyses examined the effectiveness of PPIs on well-being and depression: Sin and Lyubomirsky (2009) and Bolier et al. (2013). Sin and Lyubomirsky reported larger effects of PPIs on well-being ( Mental health has often been conceptualized as the absence of negative symptomatology . TraditiSeligman  identifiand increase well-being  with substantial heterogeneity as measured by I2 = 71.9%.The reanalysis used data reported by Sin and Lyubomirsky  in theirt(47) = 4.46, p < .001. Accordingly, we estimated the effect size after accounting for the small study size bias. The limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 50.83, p < .001. A test of residual heterogeneity indicated Q(47) = 120.24, p < .001. Thus, taking into account small study effects, the reanalyses resulted in a much smaller estimated effect size for well-being than the effect size (r = .29) reported by Sin and Lyubomirsky [bomirsky .r = .78 [95% CI = ].r = .23 [95% CI = ] with moderate heterogeneity as measured by I2 = 56.5%. t(38) = 3.19, p = .003. Accordingly, we estimated the effect size after accounting for the small study size bias. The limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 18.89, p < .001 and a test of residual heterogeneity indicated that Q(38) = 70.68, p < .001. Thus, similar to the reanalyses of Sin and Lyubomirsky’s  with substantial heterogeneity as measured by I2 = 74%.The reanalysis used data reported by Sin and Lyubomirsky  in theirt(23) = 3.20, p = .004. Accordingly, we estimated the effect size after accounting for the small study size bias. The limit meta-analysis ]. A test of small-study effects showed Q-Q'(1) = 28.40, p < .001 and a test of residual heterogeneity indicated Q(23) = 63.79, p < .001. Thus, similar to the reanalysis of well-being effect sizes, taking into account small study effects, the reanalysis of depression effect sizes resulted in a much smaller, and now non-significant estimated effect size of PPIs on depression compared to the effect size (r = .31) reported by Sin and Lyubomirsky.r = .78 [95% CI = .r = .26 [95% CI = ] with substantial heterogeneity as measured by I2 = 70.1%. t(19) = 5.33, p < .001. Accordingly, we estimated the effect size in the presence of the small study size bias. The limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 40.06, p < .001 and a test of residual heterogeneity showed Q(19) = 26.82, p = .109. Thus, similar to the re-analysis of depression effect sizes, taking into account small study effects, the replication analyses resulted in a much smaller, and now non-significant estimated effect of PPIs on depression compared to the effect size reported by Sin and Lyubomirsky (r = .31).r = .17 [95% CI = ] with moderate heterogeneity as measured by I2 = 47.1%.The reanalysis used data reported by Bolier et al.  in theirt(26) = 1.06, p = .299. Furthermore, the limit meta-analyses ], comparable to a random effect model without any adjustments. A test of small study effects showed Q-Q'(1) = 2.12, p = .145 and a test of residual heterogeneity indicated Q(26) = 48.96, p = .004. The reanalysis of Bolier et al.’s [et al.’s  subjectir = .85 [95% CI = .r = .19 [95% CI = ] with moderate heterogeneity as measured by I2 (63.1%). t(22) = 1.37, p = .184. Furthermore, the limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 4.91, p = .027 and a test of residual heterogeneity indicated Q(22) = 57.39, p < .001. These results are similar to those reported by Bolier et al. [r et al.  and obtar = .09 [95% CI = ] with heterogeneity, as measured by I2, = 35.2%.The reanalysis used data reported by Bolier et al.  in theirt(18) = 2.68, p = .02. Accordingly, it is necessary to estimate the effect size in the presence of the small study size bias. The limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 8.36, p = .004 and a test of residual heterogeneity indicated Q(18) = 20.97, p = .281.r = .06 [95% CI = ] with no heterogeneity, as measured by I2, = 0%. The regression test of the funnel plot symmetry revealed significant asymmetry t(17) = 2.13, p = .048. Accordingly, we estimated the effect size after accounting for the small study bias. The limit meta-analyses estimated an effect size of r = .01 [95% CI = ]. A test of small-study effects showed Q-Q'(1) = 3.68 p = .055 and a test of residual heterogeneity indicated Q(17) = 13.81, p = .681. Thus, a reanalysis of Bolier et al.’s [r = .10 reported by Bolier et al.The analysis was recalculated after removing one outlier (Fava.2005.1). A random effect model estimated an effect size of et al.’s  psycholor = .88 [95% CI = .r = .15 [95% CI = ] with moderate heterogeneity as measured by I2, 41.0%. t(15) = 2.66, p = . 018. Accordingly, it is necessary to estimate the effect size in the presence of the small study size bias. The limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 8.71, p = .003 and a test of residual heterogeneity indicated Q(15) = 18.41, p = .242. Thus, a replication of Bolier et al.’s [r = .10 reported by Bolier et al.et al.’s  psycholor = .10 [95% CI = ] with moderate heterogeneity as measured by I2 = 51.4%.A reanalysis used data reported by Bolier et al.  in theirt(12) = 2.71, p = .019. Accordingly, it is necessary to estimate the effect size in the presence of the small study size bias. The limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 10.14, p = .002 and a test of residual heterogeneity indicated Q(12) = 16.60, p = .165.r = .07 [95% CI = ] with some heterogeneity as measured by I2 = 27.7%. The regression test of the funnel plot symmetry revealed no significant asymmetry, t(10) = 1.55, p = .152. The limit meta-analyses estimated an effect size of r = .03 [95% CI = ]. A test of small-study effects showed Q-Q'(1) = 2.95, p = .086 and a test of residual heterogeneity indicated Q(10) = 12.27, p = .268. Thus, the reanalyses of Bolier et al.’s data revealed a smaller, non-significant effect for depression, in contrast to Bolier et al.’s finding of r = .11.The analyses were repeated after removing the outliers . A random effect model estimated an effect size of r = .81 [95% CI = . r = .14 [95% CI = ] with moderate heterogeneity as measured by I2 = 23.6%.t(12) = .52, p = .611. The limit meta-analyses ]. A test of small-study effects showed Q-Q'(1) = 0.38, p = .539 and a test of residual heterogeneity indicated Q(12) = 16.64, p = .164. However, these results are difficult to interpret due to the small number of studies.r = . 14 [95% CI = ] with no heterogeneity as measured by I2 = 0%. A regression test of the funnel plot symmetry indicated no statistically significant asymmetry, t(11) = -.17, p = .862. The limit meta-analyses estimated an effect size of r = .15 [95% CI = ]. A test of small-study effects showed Q-Q'(1) = .03, p = .862 and a test of residual heterogeneity indicated Q(11) = 11.51, p = .402. The replication analyses indicated a somewhat higher effect for depression than that reported by Bolier et al. [The effect size estimates were recalculated after the removal of an outlier (Seligman.2006.2). A random effect model estimated an effect size of r et al. .rs) originally reported by Sin and Lyubomirsky and Bolier et al. The table highlights that re-analyses of the data reported in the two previous meta-analyses resulted in much smaller effect sizes than those originally reported. Moreover, of the seven meta-analyses that yielded significant findings in the previously conducted studies, only two remained statistically significant and one more depended on one outlier when reanalyzed in the current study.The first meta-analysis examining the effectiveness of the PPIs on well-being, by Sin and Lyubomirsky , reporteAccordingly, our study had two major objectives. First, we reanalyzed the reported data from the two previous meta-analyses while taking into account study sizes and small sample size bias. Second, we replicated both meta-analyses starting with extracting relevant effect sizes directly from the primary studies rather than relying on the data published in the previous meta-analyses. In conducting these meta-analyses, the data were analyzed using a weighted random effects model while taking into account small sample size bias using the selected methods discussed above.r = .08) and depression (r = .04) than the previous authors reported . There were two major reasons for the inflated estimates reported by Sin and Lyubomirsky. First, Sin and Lyubomirsky reported effect size estimates as simple unweighted averages of study level effect sizes . This approach is inappropriate because it gives equal weight to small- and large-size studies [ail Safe N to conclude that small-study effects did not significantly inflate their findings. However, the Fail Safe N is no longer considered an appropriate way to assess small-study effects [Our reanalysis of the effect sizes reported by Sin and Lyubomirsky  revealed studies . Second, effects . The pre effects  meta-anar = .17) as reported by Bolier et al. However, the estimated effect sizes for psychological well-being (r = .02), and depression (r = .02) were smaller  than originally reported in Bolier et al. . When outliers were removed, the estimated effect sizes for psychological well-being were r = .01 and for depression were r = .07. The latter result is partially attributable to the test of funnel plot asymmetry being no longer statistically significant, in part due to the smaller number of effect sizes included. However, the limit meta-analysis estimated the effect size for depression after the removal of outliers as r = .03. The replication of Bolier et al. [r = .14. However, these results need to be viewed with caution as they are based on a small number of studies. Moreover, even though the small-study effects were not statistically significant, the number of studies was small and the scatter plots of effect sizes and study sample sizes show that large-size studies resulted in substantially smaller effects than small size studies.Our reanalysis of the effect sizes reported by Bolier et al.  revealedr et al.  meta-anar = .10 of PPIs on well-being. In contrast, the effect of PPIs on depression was nearly zero when based on the studies included in Sin and Lyubomirsky [In summary, the reanalyses and replications of Sin and Lyubomirsky  and Bolibomirsky  and highbomirsky . Notablybomirsky  includedbomirsky  in theirOur review of the two highly cited meta-analyses of PPIs resulted in a number of secondary findings and implications. First, the major reason for the larger effects reported in previous meta-analyses was that these studies did not appropriately account for prevalent small-study effects. The small-study effects are a frequent problem with meta-analyses in many fields and a number of methods  have been developed to estimate effect sizes in the presence of small-study effects. Unfortunately, these methods were not employed in the previous meta-analyses addressed by the current study. Given the presence of the small-study effects, future meta-analyses of PPIs must take into account small-study effects using appropriate estimation methods.Second, these findings are tentative because the previous meta-analyses did not include all available studies. To illustrate, Bolier et al.’s  inclusioThird, the failure to include all available studies in the previous meta-analyses suggests the need for a comprehensive meta-analysis of PPIs effect on well-being starting with a comprehensive search for relevant studies. A preliminary search using PsycInfo for studies of PPIs using only the most obvious search strategy  yielded over 200 relevant studies, more than tripling the number of studies included in the previous meta-analyses.Fourth, our review of the primary studies included in the previous meta-analyses revealed persistent limitations with their method and results sections. In general, no primary studies with pre-post designs reported pre-post correlations for outcome measures, which are necessary to calculate the most appropriate effect sizes . Though Fifth, it is evident from the diverse inclusion and exclusion criteria of previous meta-analyses that there is no consensus as to what constitutes a PPI. Bolier et al.  excludedThus, the “true” effects of PPIs may be substantially different from what Sin and Lyubomirsky and Bolier et al. meta-analyses indicate. While our re-analyses and replications of these meta-analyses converge and indicate that the effect of the PPIs are relatively small when small sample bias is taken into account, estimates of effect sizes are not definitive because neither Sin and Lyubomirsky nor Bolier et al. meta-analyses were comprehensive and a large number of relevant studies are likely missing.Accordingly, a comprehensive and transparent meta-analysis of all relevant studies of PPIs is necessary and is likely to have a major influence on the field. Such a meta-analysis is likely to allow for meaningful moderator analyses in answering questions such as: Is group administration more effective than individual administration? Are longer interventions more effective than shorter interventions? Are some types of interventions more effective than other types of interventions? Importantly, a comprehensive meta-analysis is likely to provide a more definitive determination of how effective PPIs are at increasing well-being.Given that our meta-analyses indicate that the effects of the PPIs on well-being and depression may be smaller than previously reported, future research may need to employ strategies likely to increase the effectiveness of PPIs. For example, PPIs are likely to be more effective if they are deployed over longer periods of time . Some reThe current study re-analyzed the data reported in previous meta-analyses that examined the effectiveness of PPIs on increasing well-being and decreasing depression, as well as completely replicated  the previous meta-analyses. The reanalysis of the previously reported data showed that although correlations between the recalculated effect sizes and the previous meta-analyses effect sizes were fairly high (suggesting that the same data were extracted), the effect sizes were lower than previously reported and often nonsignificant. The major contributing factor for this discrepancy was that the present study accounted for the strong presence of small-sample size bias. Critically, both meta-analyses reviewed, did not include a large number of relevant studies, and thus, effect sizes estimated from their sample of primary studies need to be confirmed by future, more comprehensive, meta-analyses. Accordingly, a comprehensive and transparent meta-analysis of all relevant studies of PPIs is necessary. Such a meta-analysis will allow for meaningful moderator analyses to determine effects of various PPIs including whether individual PPIs are more effective than group PPIs and whether longer and more intense PPIs are more effective than shorter and less intense interventions. Our research underscores that any future meta-analyses of PPI effectiveness ought to take into account frequent methodological issues such as prevalent small sample size bias.


Diverticulosis and colorectal neoplasia share epidemiological trends and risk factors which are common in Western countries and incidences increase with age. However, the data on an association between diverticulosis and colorectal neoplasia are conflicting. Thus, we performed a meta-analysis to evaluate whether diverticulosis is associated with colorectal neoplasia.A systematic literature search of PubMed, EMBASE, Cochrane Library, Web of Science, and SCOPUS was conducted to identify studies that investigated the association between diverticulosis and advanced colorectal neoplasia , adenomas, or polyps. The demographic characteristics of patients, including age, gender, indication for colonoscopy, confounding factors, and outcomes of colorectal neoplasia were assessed.We identified 29 cross-sectional studies  that investigated the association between diverticulosis and colorectal neoplasia. The meta-analysis found that diverticulosis was not associated with advanced colorectal neoplasia . Although there was a positive correlation between diverticulosis and adenomas  and diverticulosis and polyps , diverticulosis did not increase the risk of adenomas  in patients who underwent screening colonoscopy. Moreover, all the increased risk of colorectal neoplasia in patients with diverticulosis was observed in published studies only, and not in unpublished ones.This meta-analysis demonstrated that diverticulosis is not associated with an increased risk of advanced colorectal neoplasia. Although diverticulosis was associated with a higher risk of polyps and adenomas, the risk was not increased in screening populations. Moreover, the increased risk of colorectal neoplasia in patients with diverticulosis was observed only in published studies and not in unpublished ones. Colonic diverticulosis is the condition of outpouching of the colonic mucosa and submucosa as a result of weakness in the muscle layers of the colonic wall. Patients with diverticulosis are usually asymptomatic; only 15%~25% of patients with diverticulosis develop complications such as diverticulitis and gastrointestinal bleeding . SimilarDiverticulosis and colorectal neoplasia share risk factors that are common to Western countries, such as smoking, alcohol, constipation, and low fiber diet , 5, 6. TWe used multiple comprehensive databases to identify literature investigating the association between diverticulosis and advanced colorectal neoplasia , adenomas and polyps. This study is based on the Cochrane Review Methods .We searched MEDLINE (1964 to Jan 2019), EMBASE (1947 to Jan 2019), the Cochrane Central Register of Controlled Trials (1966 to Jan 2019), Web of Science (1964 to Jan 2019), and SCOPUS (1964 to Jan 2019). We did not restrict our search with regard to language or year of publication. The following keywords and medical subject headings (MeSH) were searched via MEDLINE:  AND . See The inclusion of all studies was independently decided by two reviewers (HJL and HJK) based on the selection criteria. Study selection was performed through 2 levels of screening: At the first level, we screened titles and abstracts of identified studies. At the second level, we screened the full text. Studies were included in our meta-analysis if they satisfied the following: (1) diverticulosis documented by colonoscopic examination; (2) the outcomes reported including advanced colorectal neoplasia , adenomas, and polyps; (3) relative risks (RR) or odds ratio (OR) reported or data provided for their calculation; and (4) were cohort, cross-sectional and case-control studies, but neither case reports nor reviews. Neither type of publication nor language were restricted. Articles were excluded from this meta-analysis if the published data were insufficient for estimating the relative risks/odds ratios (RRs/ORs) and 95% confidence intervals (CIs).The two reviewers independently extracted data from each study using a predefined data extraction form. Any disagreement unresolved by discussion was then reviewed by the third author (SJP). The following variables were extracted from studies: the first author; journal name; year of publication; study design; country/ethnicity of study participants; publication type; study period; total number of patients; age; gender; indication for colonoscopy (screening vs. diagnosis of symptoms); confounding factors; outcomes including the presence of advanced colorectal neoplasia , adenomas, and polyps; and RR/OR and 95% CIs. Diverticulosis was defined as the presence of endoscopically diagnosed diverticuli in any part of the colon. Advanced colorectal neoplasia consisted of advanced adenoma and/or CRC. An advanced adenoma was defined as a tubular adenoma ≥1cm in size, any polypoid lesion with a villous histology, high grade dysplasia. A polyp was defined as any localized projection above the surrounding colonic mucosa regardless of histologic evaluation. Colonic segments proximal to the splenic flexure, which included the cecum, ascending, and transverse colon, were defined as proximal colon; and portions of the colon distal to the splenic flexure, including the descending and sigmoid colon, and rectum, as distal colon. If any of the variables listed in this section were not mentioned in a study, we requested the data via email.Two reviewers (HJL and HJK) independently assessed the methodological quality of each study, using the appraisal tool for cross-sectional studies (AXIS tool) . Any unrI2 statistic, with values of 25%, 50%, and 75% considered low, moderate, and high, respectively. If the data were available, we used effect sizes from multivariate models, with confounding factors adjusted in each study. We performed subgroup analysis to assess the association between diverticulosis and adenomas and polyps according to the specific indication for colonoscopy (screening vs. diagnosis of symptoms).The primary outcome of our review was to determine the risk of advanced colorectal neoplasia (advanced adenoma and/or CRC) based on the presence of diverticulosis. As a secondary outcome, we also evaluated the associations between diverticulosis and adenomas and polyps. We conducted pooled analyses using the inverse variance method, with random-effects weighing for meta-analyses of outcomes reported by multiple studies that were sufficiently similar to justify combining results. To estimate heterogeneity, we estimated the proportion between-study inconsistency due to true differences between studies (rather than differences due to random error or chance) using We performed sensitivity analysis after excluding studies with low methodological quality or abstract-only studies to check whether the results had changed. If the results did not change significantly after excluding low-quality studies (abstract-only study), then they were considered to be robust. If the results had changed or the conclusions differed, then they had low stability.P < 0.05 suggesting a significant publication bias. We used RevMan version 5.2  and Stata 13.1/MP version for these analyses. An OR of >1 favored the risk factor and P < 0.05 and a 95% CI that did not include the value 1 were considered to be statistically significant.The Egger test was used to evaluate publication bias, with A flow diagram of our review is shown in A total of 29 cross-sectional studies  investigated the association between diverticulosis and colorectal neoplasia. We used a standardized protocol and report form to abstract the following data from each publication: first author, year of publication, country/ethnicity, publication type, study period, total number of patients, prevalence of diverticulosis (%), mean age, male (%), confounding factors, and reported outcomes. The studies were published between 2000 and 2019, with an enrollment period ranging from 1995 to 2018. Among the 450,953 participants analyzed in this meta-analysis, 32,235  had diverticulosis, 53.4% were male, and the mean age was 56.0 years. In addition, 19 studies were performed in Western countries and 9 studies in Asia. Of these 29 studies, 16 reported data on advanced colorectal neoplasia, 19 on adenomas, and 14 on polyps. The characteristics of these studies are summarized in The quality assessments of the individual studies are listed in I2 = 96%)  .I2 = 90%) (I2 = 100%) (I2 = 65%) and polyps   . Fourtee = 100%) . Multiva2 = 57%)  Figs.I2 = 94%) (I2 = 98%)  . Similar2 = 98%) .When focusing on serrated polyps, 3 studies reported the association between diverticulosis and serrated polyps , 13, 25.I2 = 75.8%) (I2 = 44.2%) and polyps   . Statist= 82.5%) .I2 = 76.4%)  .According to the Egger test, there was no evidence of funnel plots asymmetry between the diverticulosis and advanced colorectal neoplasia, adenomas, and polyps Figs .In this meta-analysis, we combined evidence from 29 cross-sectional studies involving 450,953 patients to investigate the association between diverticulosis and colorectal neoplasia. We found that diverticulosis was not associated with advanced colorectal neoplasia. Though diverticulosis was associated with a higher risk of adenomas and polyps, the risk of adenomas was not increased in patients with diverticulosis who underwent screening colonoscopy. Moreover, the increased risk of colorectal neoplasia in patients with diverticulosis was observed only in published studies and not in unpublished ones.The pathologic mechanisms involved in both diverticulosis and colorectal neoplasia have been found to include diets low in dietary fiber and rich in saturated fat, and slow colonic transit times , 8, 41. In 2008, a systematic review was performed to investigate the possible association between diverticulosis and colorectal neoplasia . AlthougOur meta-analysis demonstrated that diverticulosis was not associated with an increased risk of advanced neoplasia. It is in agreement with two nationwide population-based studies reporting that except for the first year after the diagnosis of diverticular disease, diverticular disease did not appear to be associated with an increased risk of subsequent CRC , 50. TheIn contrast, our data revealed that patients with diverticulosis had a 1.47-fold risk of colorectal adenomas and a 1.95-fold increased risk of polyps compared to controls without diverticulosis. Indeed, many studies reported a positive correlation between diverticulosis and adenomas , 44, 51 It is noteworthy that the associations between diverticulosis and colorectal neoplasia were clearly varied, based on the type of publication. Diverticulosis was associated with an increased risk of adenomas and polyps in published studies, but no association was found in unpublished data. Studies of patients who underwent colonoscopy due to gastrointestinal symptoms were all published, in contrast, regarding screening colonoscopy, the risk were obviously different according to publication types. A recent meta-analysis pointed out their drawback of publication bias because funnel plots are asymmetric that could interfere with the interpretation of results . Our revOur study has several limitations. First, since all the included studies in our meta-analysis were cross-sectional studies, the causal relationship between diverticulosis and colorectal neoplasia cannot be assessed. Second, adenoma represented the principal precursor to CRC , but alsIn conclusion, our meta-analysis demonstrated that diverticulosis was not associated with an increased risk of advanced colorectal neoplasia. Although diverticulosis is associated with a higher risk of adenomas and polyps, the increased risk is only observed in patients with gastrointestinal symptoms and not in screening populations. In addition, the effect of diverticulosis on colorectal neoplasia in currently available studies might be overestimated to get a conclusion. A prospective cohort study to elucidate not only the association of diverticulosis with colorectal neoplasia but also the causality of the two diseases is warranted.S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.S4 Fig(TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S1 Appendix(DOCX)Click here for additional data file.


Background: Vitamin-D binding protein (DBP) also known as GC protein, is a major determinant for vitamin- D metabolism and transport. GC1F, GC1S, and GC2 are the three allelic variants (denoted as rs4588 and rs7041) of GC, and known to be associated with chronic obstructive pulmonary disease (COPD). However, contradictory reports and population specific risk attributed by these alleles warranted detailed genetic epidemiology study to establish the association between GC variants and COPD. In this study we performed a meta-analysis and investigated the genetic architecture of GC locus to establish the association and uncover the plausible reason for allelic heterogeneity.Methods: Published cross-sectional case control studies were screened and meta-analysis was performed between GC variants and COPD outcome. RevMan-v5.3 software was used to perform random and/or fixed models to calculate pooled odds ratio (Meta-OR). Linkage disequilibrium (LD) and haplotypes at GC locus were evaluated using 1000 Genomes genotype data. In silico functional implications of rs4588 and rs7041 was tested using publicly available tools.Results: GC1F allele and GC1F/1F genotype were found to confer COPD risk in overall meta-analysis. GC1S/1S was found to confer risk only among Europeans. In silico investigation of rs4588 and rs7041 identified strong eQTL effects and potential role in regulation of GC expression. Large differences in allele frequencies, linkage disequilibrium (LD) and haplotypes were identified at GC locus across different populations , which may explain the variable association of different GC alleles in different populations.Conclusion: GC1F and GC1F/1F impose significant genetic risk for COPD, among Asians. Considerable differences in allele frequencies and LD structure in GC locus may impose population specific risk. Chronic obstructive pulmonary disease (COPD) is a complex disease affecting the lung function. Genetically susceptible individuals develop the COPD while they get exposed to environmental triggers, such as noxious gases or suspended particles. Decreased level of vitamin-D in serum is associated with COPD among individuals with a history of smoking  and other diseases including COPD D3, and thus affect its serum concentration D3 by catabolizing enzymes. More than 90% of the circulating 25(OH)D3 present in tightly bound (Kd ~ 10−9 M) form with GC proteins , also known as group-specific component (GC), belongs to a gene cluster family which is expressed in liver and other tissues D3 was observed in the ascending order of GC genotypes 1F/1F <1S/1F <1S/1S <1S/2 <2/2 among COPD patients , which are responsible for the emphysema degradation of lung alveoli. Thus, optimal serum concentration of Vitamin-D3 is very critical among emphysema patients and a trial for such serum Vitamin D3 intervention among a large participant group can further elucidate its role in COPD progression  in a well-defined COPD cohort  is also involved in the inflammation by getting converted into MAF (Macrophage Activating Factor) in the presence of enzymes secreted by leucocytes. It has been found that the conversion of GC into GC-MAF is a deglycosylation process. Absence of glycosylated GC alleles with COPD and investigate the genetic and functional aspects of GC alleles. Locus architecture of different populations was also investigated to explain the non-replication/differential replication of GC alleles in different populations. We hypothesized that genetics architecture at GC locus leads to population specific allelic variation in GC and its association with COPD. The study was performed with the following specific objectives: (i) perform meta-analysis to establish association of commonly studied GC alleles with COPD, and (ii) investigate the genetic heterogeneity at a functionally relevant GC locus, that explain variability in GC protein and COPD.It is evident that GC is a major determinant for several health parameters including those associated with COPD. However, contradictory findings of association of different alleles with COPD and non-replication across different populations warranted further meta-analysis and detailed population genetics studies. In the present study, we anticipated to explain the association of known GC has been tested with COPD. We restricted our study to three major genetic polymorphisms of GC, namely GC1F, GC1S, and GC2 alleles. Literature was searched online in the National Center for Biotechnology Information (NCBI-PubMed), Google Scholar and Medline. The major search language for the literature was English, papers in other languages were translated for further review. To obtain the best quality outcome, we include only peer reviewed scientific literature. Literature were searched until May 2018. The keywords used for the search for literature were as follows: Vitamin D binding protein and chronic obstructive pulmonary disease, DBP and COPD, GC alleles and COPD, COPD association GC. Cross references were also reviewed and references from the retrieved articles were also checked manually so as to find any relevant articles.Our objective was to identify research articles where genetic association of GC were studied for their association with COPD. Included studies clearly mentioned either the actual numbers, or the percentage of cases and controls with different genotypes and alleles of GC. Included studies have both smokers and non-smokers among both cases and controls.Only case-control studies were included for this meta-analysis. Only those studies were included where different alleles  and genotypes  of Data was extracted from eligible articles by two investigators independently and differences and controversies were resolved by group discussions. We first validated the study types and then extracted author names, year of publication, details of genotypes/alleles and their frequencies in COPD patients and controls.GC1F, GC1S and GC2, represented in NCBI dbSNP as rs4588 and rs7041, respectively  Copenhagen: The Nordic Cochrane Center, The Cochrane Collaboration, 2014. Additive genetic model with 95% confidence interval (CI) was used in each of these independent analyses. Heterogeneity between studies was calculated by the I2 and chi2 test, where I2> 50% and chi2p < 0.05 was considered as significant heterogeneity. Meta-analysis of odds ratios were performed using a random effect model where significant heterogeneity was observed, otherwise a fixed effect model was used. Overall effect size (Meta-OR) was calculated by Z-test with 5% alpha level. A sensitivity analysis was performed to access whether meta-analysis results were substantially influenced by the presence of any study. This was done by systematically excluding one study at a time and recalculating the significance  of the results. The funnel plot was used to analyze the publication bias. Subgroup analysis between Asian and Caucasian studies was also performed to identify any significant differences due to individual group stratification.Results of association of three distinct alleles have been included in this study. These alleles were ectively . TherefoGC locus was evaluated to explain population specific effects (if any) of GC alleles on its association with different human traits/diseases. To analyze the linkage disequilibrium, LD plots and haplotypes were reconstructed using Haploview , HaploReg .Genetic architecture of http://www.regulomedb.org/index) was used to analyze the regulatory function and GTEx portal  was used to analyze single tissue or gene eQTL.Functional implications of rs4588 and rs7041 were analyzed using open source browsers. RegulomeDB (n = 5), which was done previously on COPD and GC, was also excluded but was used to identify cross references. Fourteen studies were excluded because they were either cohort studies or random clinical trials done on supplementation of vitamin-D3. Eleven studies were found irrelevant, either due to less information for cases or control subjects, and one study was in other language and was excluded from the meta-analysis. A further seven studies were excluded as adequate/complete genotype and study participant information were not given. After this screening based on our inclusion/exclusion criteria, a total of 14 studies were found eligible for meta-analysis   . Meta-anGC1F allele has been found significantly predisposing for COPD outcome in combined analysis . Independently, GC1F allele has been found strongly associated among Asians , but not among Europeans   . Both GC outcome . HoweverGC1F/1F was found significantly predisposing genotype with COPD outcome . Independent analysis found significant association of this genotype among Asians , but it remains insignificant among Europeans . Significant predisposition was observed for GC1S/1S genotype among Europeans , however it remains insignificant among Asians  was observed. Subgroup analyses did not identify any significant (p<0.05) subgroup stratification  and GIH , however, these haplotypes were not found in JPT and YRI. Furthermore, moderate yet similar LD was observed between these two markers in CEU  and GIH , however, LD is completely broken in JPT  and YRI . Notable haplotypic variations were observed across the genomic region, whereas in JPT and YRI, these two variations are not in tight linkage with neighboring markers (r2 = 0) was observed among different African populations, whereas the highest degree of LD was observed among Europeans and South Asian populations followed by Americans  expresses in the liver despite very negligible expression in the pancreas and stomach. For two missense SNPs, rs4588, and rs7041, no evidence was observed for significant eQTL on GC in liver tissue, however, significant eQTL was observed in subcutaneous adipose (p = 6.55E-6), sun exposed skin (p = 1.67E-6), and stomach (p = 5.46E-9) tissues. SNP rs4588 was identified to alter motif-binding sites of transcription factors SP1 and SP3; and transcription factor binding element (KLF16). rs4588 and rs7041 were both identified: (a) to localize in DNase hypersensitivity regions in a common set of cell types and tissues, and (b) potentially alter histone modification in liver (strongly) and skin (quiescent/low) tissue.GC polymorphisms with COPD. This meta-analysis established that GC1F allele and GC1F/1F genotype confers risk of COPD. However, association is majorly restricted to Asians and not in Europeans. On the contrary, GC1S/1S genotype was observed to confer risk to Europeans only (with borderline significance). At least one copy of GC2 has been found to confer protection from COPD among both Asians and Europeans. Previous meta-analysis studies and independent reports have shown different results from Europeans and Asians, which could be due to differences in allelic segregation and haplotypic heterogeneity at a population level  from African countries, we have included their representative genotypes for comparative genetic studies (GC variants (3 (Braun et al., GC, and thus could play critical role in regulating expression and function of the GC protein.In this systematic review, we performed meta-analysis and evaluated linkage disequilibrium at GC locus, in order to investigate the association of common  studies . Notablevariants . These m3 is reported to be associated with the severity of COPD condition. Particularly, rs4588 has been shown to influence GC binding to Vitamin-D3 (Nimitphong et al., 3 intake/supplementation, a functionally more potent form of GC is necessary to maintain optimal serum bioavailability of vitamin-D3. Therefore, inter individual differences in the GC protein may act as a predisposing factor for COPD. Further genetic epidemiological studies are warranted to identify novel risk alleles from GC that are associated with GC function, and thus implication in COPD. However, the presence of differential LD structure of GC locus needs to be considered as a major confounding factor.Although VDD is found to be associated with COPD (Jolliffe et al., in silico genetic study. RK, DN, and SS contributed in writing the manuscript and interpreted the results. All the authors reviewed the manuscript and finalized for submission.SS conceptualized and designed the study. RK and DN performed literature screening and meta-analysis. SS performed The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Several researchers have suggested that the rs243865 (16q13-q21) polymorphism in the promoter region of the metalloproteinase-2 (MMP-2) gene could be associated with an increased risk of developing age-related macular degeneration (AMD). However, previous results remain inconclusive. To clarify this controversy, we conducted a meta-analysis of the relationship between rs243865 of MMP-2 and AMD.We included 6 independent case-control studies involving 1,682 AMD patients and 2,295 healthy subjects. The association between the rs243865 polymorphism and AMD was examined by the overall odds ratio (OR) with a 95% confidence interval (CI). We used a recessive genetic model analysis, sensitivity analysis, and assessment of bias in our meta-analysis.Our results showed that there was no significant association between the variant T allele  or the CT+TT genotype  of rs243865 MMP-2 polymorphism and the presence of AMD.The rs243865 MMP-2 polymorphism was not associated with an increased risk of developing AMD. The MMP-2 (-1306 C>T) promoter variant is unlikely to have a major role in AMD risk susceptibility. Age-related macular degeneration (AMD) is the leading cause of severe vision loss and blindness worldwide, and the incidence in developed countries ranges from 9% to 25% at ages 65 to 75 years ,2. The rThe development of AMD is influenced by lifestyle (smoking), dietary factors, drusen accumulation, and parainflammation. In addition, the incidence of patients with AMD is also affected by genetic components ,6. CurreThe current study conformed to the checklist for meta-analysis of genetic association studies specified for PLOS One approach  16]..16].To identify eligible studies, we searched in the PubMed, Web of Science, Scopus, and Embase electronic databases. Potentially relevant articles were sought by using the search terms in combination as Medical Subject Headings (MeSH) terms and text words: “matrix metalloproteinase-2”, “polymorphism”, “rs243865”, “-1306C>T”, “age-related macular degeneration”, “AMD”, “macular degeneration”, and “maculopathy”. We also scanned the reference lists of the retrieved publications to identify additional relevant articles .The following inclusion criteria were used to identify published articles for our meta-analysis: (1) associations between the rs243865 MMP-2 polymorphism and AMD published in the last 18 years; (2) study populations included only individuals not related to one another; (3) sufficient genotype data were presented to allow calculation of the odds ratios (ORs); (4) each study clearly described the diagnosis of AMD, which was limited to the National Institute for Health and Care Excellence (NICE) Age-Related Maculopathy grading system or the International Classification and Grading system ,18; (5) To improve the reliability of our results, two investigators  used a standardised form to independently extract data. Any divergences of opinion were resolved by consensus with the senior authors . Second and third literature searches were conducted but did not contribute additional studies for the analyses. Citations were first scanned at the title/abstract level. Shortlisted studies were then retrieved in full text.From each study, the following information was extracted: the author´s name, year of publication, location, and ethnicity and demographic information (age and sex) of the study population. Allele and genotype frequencies were extracted or calculated. HWE was verified for each eligible study. The corresponding authors of the original studies were contacted if further data were needed.Study quality was ranked as high, moderate, or low  by the Newcastle-Ottawa Quality Assessment Scale (NOS) . Two indThe main meta-analysis compared the presence of the variant T allele of rs243865 MMP-2 polymorphism among AMD patients as cases versus healthy subjects as controls. If three or more studies were available, further meta-analyses were performed comparing: (1) the presence of CT+TT genotype of rs243865 MMP-2 polymorphism among AMD patients and healthy subjects; (2) the presence of CT+TT genotype of rs243865 MMP-2 polymorphism among AMD patients and healthy subjects by age; and (3) the presence of CT+TT genotype of rs243865 MMP-2 polymorphism among AMD patients > 65 years old and AMD patients < 65 years old.2 statistic was determined, representing the percentage of the observed study variability attributed to heterogeneity rather than to chance [2 test.For each study, we used a random-effects model  to calcuo chance . Begger´o chance . DeviatiA flow diagram of the study selection process is shown in Each of the articles analysed in our meta-analysis included the genotypic and allelic frequencies of rs243865 MMP-2 polymorphism in AMD patients and healthy controls . All of After analysis of the pooled selected studies, there were no significant associations between the variant T allele . There wx-axis and the inverse of variance of the effect on the y-axis Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOCX)Click here for additional data file.


Tendons and ligaments are common sites of musculoskeletal injuries especially during physical activity. The multifactorial etiology of tendon-ligament injury (TLI) includes both genetic and environmental factors. The genetic component could render influence on TLI risk to be either elevation or reduction.COL5A1) polymorphisms, mainly rs12722 (BstUI) and rs13946 (DpnII), with TLI warrant a meta-analysis to determine more precise pooled associations.Inconsistency of reported associations of the collagen type V alpha 1 chain (Multi-database literature search yielded eight articles (11 studies) for inclusion. Pooled odds ratios (ORs) and 95% confidence intervals were used to estimate associations. Heterogeneity of outcomes warranted examining their sources with outlier treatment.p = 0.0009–0.04) in the pre-outlier analysis were non-heterogeneous (p > 0.10). The non-significant and heterogeneous  pre-outlier rs12722 and rs13946 results became significant  and heterogeneity eliminated (I2 = 0%) with outlier treatment. Significant associations  were also observed in other COL5A1 polymorphisms (rs71746744 and rs16399). Sensitivity analysis deemed all significant outcomes to be robust.All rs12722 effects indicated reduced risk (OR < 1.0). The significant outcomes  contains supplementary material, which is available to authorized users. COL5A1 polymorphisms reduce the risk of tendon-ligament injury among Caucasians.Pre-outlier reduced risk effects on tendon-ligament injury were significant in rs12722 but not in rs13946.Outlier treatment impacted upon the rs12722 and rs13946 polymorphism outcomes, with significance strengthened and gained, respectively.Non-heterogeneous significant reduced risk of tendon injury observed in rs71746744 and rs16399 obviated the need for outlier treatment.COL5A1) gene, located on the long (q) arm of chromosome 9 [COL5A1 gene have been found to impact upon TLI, the most prominent being rs12722 (BstUI) and rs13946 (DpnII) which are found within the 3′-untranslated region (UTR) [COL5A1 polymorphisms  have been considered as risk modifiers [COL5A1 polymorphisms in TLI, but results have been inconsistent. Inconsistency of results may be attributed to lack of statistical power because of small sample sizes. Meta-analysis synthesizes primary study data yielding an aggregate sample size to indicate raised statistical power. Therefore, we perform a meta-analysis of all available data on COL5A1 polymorphisms and their relationship with TLI so that we could obtain better estimates of associations.Normal tendons and ligaments differ in function and are impacted under conditions of injury . Tendon-mosome 9  and is eon (UTR) . Other CCOL5A1,” “ligament injury,” “tendon injury,” and “polymorphism” as medical subject heading and text, restricted to English. References cited in the retrieved articles were also screened manually to identify additional eligible studies. Inclusion criteria were (i) case–control studies evaluating the association between COL5A1 polymorphisms and risk for TLI, (ii) sufficient genotype/allele frequency data presented to calculate the odds ratios (ORs) and 95% confidence intervals (CIs), and (iii) participants were either athletes or non-athletes. Exclusion criteria were (i) non-English articles and (ii) studies whose genotype or allele frequencies were unavailable or, when they are, combined with other polymorphisms, preventing proper data extraction.We searched MEDLINE using PubMed, Science Direct, and Google Scholar for association studies as of April 15, 2018. The terms used were “type V collagen,” “COL5A1 polymorphism type, and basis for matching (Table χ2 test.Two investigators (NP and PT) independently extracted data and arrived at a consensus. Extracted data were tabulated; when needed, we contacted authors of the original articles to request for additional information. The following information were obtained from each publication: first author’s name, published year, country of origin, ethnicity, TLI and p values for HWE. Confining the analyses to HWE-compliant studies constituted modifier treatment. Subgrouping was based on injury type (tendon or ligament).Additional file p values, power, corrections for multiplicity, comparative sample sizes between cases and controls, use of the HWE, and genotyping methods. CB scores range from 0 (worst) to 10 (best) where scoring is based on quality .We used the Clark–Baudouin (CB) scale to evaluate methodological quality of the included studies . We founCOL5A1 polymorphisms  warranted the use of a common notation indicating variant (var) and wild-type (wt) alleles. Additional file COL5A1 polymorphisms. After estimating TLI risk (OR) for each study, pooled ORs with 95% CIs were calculated for the following genetic models: (i) homozygous: (var–var and wt–wt) genotypes compared with wt–wt, (ii) recessive: (var–var versus wt–var+wt–wt), (iii) dominant: (wt–wt versus wt–var+var–var), and (iv) codominant: (var versus wt). To compare effects on the same baseline, we used raw data for genotype frequencies to calculate pooled ORs, using either fixed [χ2-based Q test [I2 statistic which measures degree of inconsistency among studies [p values of < 0.05 were considered significant except in estimations of heterogeneity. Given the low power of the χ2-based Q test for heterogeneity, the p value was set at < 0.10 [Examining five er fixed  or randod Q test , (ii) qu studies , and  and an α level of 0.05 (two-tail), statistical powers are adequate for rs12722 (99%) and rs13946 (96%). Two studies [A1 and A2 alleles which are BstUI products of restriction fragment length polymorphism due to two single-nucleotide polymorphism substitutions, the A2 allele referring to the variant C allele [p values for HWE (up to 10−5). We minimized this by combining the A1 and A2 data. Three studies from two articles were HWE-non-compliant [Table  studies , 28 repoC allele . Extractompliant , 31. Thep < 0.05) indicating reduced (protective) risk (OR < 1.0) were observed in all significant ORs and non-significant ORs that were altered with outlier treatment as well as all post-outlier outcomes (Table Pooled effects (es Table . All posCOL5A1 with TLI. Table p = 0.003–0.04) and modifier  analyses. Non-significant heterogeneous pooled effects in rs12722 and rs13946  were altered to significance with outlier treatment, outcomes of which are summarized in Table p = 0.0001–0.0003) and rs13946 , modifier (HWE) outcomes in rs12722 .Our hypothesis-driven approach indicates associations of p = 0.49) and heterogeneous (I2 = 69%). Sources of this heterogeneity were identified with the Galbraith plot visualized in Fig. p = 0.0003) and eliminated heterogeneity (I2 = 0%).Operation of outlier treatment is visualized in Figs. p = 0.0009–0.003). Non-significant tendon subgroup outcomes  were altered to significance  following outlier treatment (Table Table p = 0.003–0.03) and rs16399  with tendon injuries only but not rs319378 . These associations were either non-heterogeneous (I2 = 12–46%) or homogeneous (I2 = 0%). Table pheterogeneity < 0.10). Outlier treatment impacts upon heterogeneity by reducing (pheterogeneity > 0.10) or eliminating it (I2 = 0%). Table I2 = 0%). Table I2 = 0%).Pre-outlier ORs were generally heterogeneous .Table p < 0.05) than rs13946. For both polymorphisms, the most associated models with TLI were homozygous and recessive more than dominant/codominant. On the whole, this meta-analysis delineated which polymorphisms in the COL5A1 gene have associations with TLI  and those that do not (rs319378) as well as those that were altered (rs12722 and rs13946) with meta-analytical treatment  such that significance was either intensified or gained. The main finding of this study points to significant reduced risk effects , up to 41% and 55%, pre- and post-outlier, respectively, in rs12722 and up to 68% in rs13946, post-outlier only. Outlier treatment impacted upon heterogeneity, significance, and precision of outcomes, seen in the overall, modifier, and subgroup analyses. Significant effects (up to 69%) of the other COL5A1 polymorphisms were consistent and homogeneous (most had I2 = 0%) indicating associations of rs71746744 and rs16399 with tendon injury. The combination of significance, consistency, homogeneity, and increased precision of pooled effects in these comparisons improved our findings.Applying meta-analytical techniques is methodologically complex. This complexity arises from interpreting outcomes resulting from the combined applications of genetic modeling, modifier, outlier, and subgroup analyses as well sensitivity treatment. These considered, rs12722 was more significant  and COL5A1 but did not differentiate genotype data between these two genes, which was our reason for excluding it in our meta-analysis. Of note, MMP3 was genotyped but not COL5A1. Second, we had two added studies [TT versus TC+CC for the previous study [var–var versus wt–var+wt–wt in our study. Given this contrast, calculations of the ORs would inevitably have taken diverging outcomes. Thus, OR risks were increased for the previous study [We compare our findings with a recent (March 2018) meta-analysis  which exv et al.  and oursv et al. –30, 32. v et al.  includedv et al.  who exam studies , 31 whic studies . Third, us study  and var–us study  and reduus study ; (ii) thus study .CC genotype: (i) decreased AT risk of up to 62% in different populations  [A2 and A1 alleles in rs12722 which elicited protective and increased risks in two studies, respectively [CC genotype had a significantly decreased risk of developing AT compared with those with a T allele in either TT or TC genotypes [Comparing our reduced risk pooled findings with the component study-specific ORs shows the following for the rs12722  Africa) , (ii) re Africa) , 30, 32,ectively , 28. In enotypes .COL5A1 gene 3′-UTR region affect mRNA stability and its export from the nucleus after transcription where regulatory sequences control gene expression at the posttranscriptional level [COL5A1 mRNA [COL5A1 gene where rs71746744 and rs16399 were found to have significant association with AT. The rs71746744 variant del/del genotype was found to be associated with reduced risk of AT. Furthermore, they surmised linkage of rs12722 with rs71746744 and rs16399 [Genetic variations in the al level . Therefoal level . Functio5A1 mRNA . Abraham5A1 mRNA  investig rs16399  which suCOL5A1 polymorphisms ranged from presence to absence of associations. In their presence, risk effects for the variant genotype were increased or reduced, significant or not. Meta-analysis, however, gives more information in reporting effects for COL5A. These involve exploring magnitude and precision of effects, as well as consistency and stability, all in consideration of heterogeneous outcomes. These features raise the levels of evidence to support conclusions on the associations of COL5A1 polymorphisms with TLI.Reported findings on the role of genetic variants in TLI have differed between studies. Several methodological problems may explain the discrepancies, including limited statistical power, unrecognized confounding factors, misleading definition of phenotypes, and stratification of populations . ReportiCC genotype among ACLR participants; (ii) A1 and A2 alleles for BstUI were not examined separately, but instead, we combined them which may have sacrificed precision of outcomes, given the reported contrasting effects of these two alleles in the literature [COL5A1 polymorphisms in the component studies [n = 2–3) and underpowered status (58% and 44%) warrant caution in interpreting the significant effects of rs71746744 and rs16399.Interpreting our findings should be contextualized in view of its strengths and limitations. Limitations of our study include the following: (i) we did not examine gender effects due to insufficiency of data. Nevertheless, Posthumus et al.  examinedterature , 28;  confining our meta-analysis to Caucasians rendered epidemiological homogeneity to the study which thus excludes potential confounding effects of population stratification; (ii) non-HWE-compliant studies were a minority which minimizes the issue of genotyping errors thus avoiding methodological weaknesses in the summary outputs . BesidesCOL5A1 polymorphisms with TLI are uncovered. Genetic structure of the homozygous and recessive models point to the variant allele as protectively associated with TLI. TLI is a complex condition involving interactions of several genetic and non-genetic risk factors. Gene-gene and gene-environment interactions have been reported to have roles in the associations of COL5A1 polymorphisms with TLI [COL5A1 polymorphisms and TLI. Injury-related issues in sports medicine beg comprehensive investigation of its risks. Our contribution is the synthesis approach offered by meta-analysis in elevating the level of evidence. Given the focus of this study, we hope to have clarified the genetic risks posed by COL5A1 polymorphisms to TLI. The mainly protective findings of COL5A1 rs12722 and rs13946 polymorphisms may be modest given our focus on just one gene. However, the evidence we present hopes to contribute to better understanding of the genetic nature of TLI.The importance of our results is underpinned by the fact that each component study in this meta-analysis lacked adequate statistical power, but when combined using meta-analysis, clear reduced risk associations of with TLI . None ofwith TLI , 31, 32.Additional file 1:Table S1. Quantitative characteristics of the included studies that examined the association of COL5A1 polymorphisms with TLI studies. (DOCX 57 kb)Additional file 2:Table S2. PRISMA checklist. (DOCX 30 kb)


The summary RR was 1.16  for participants with diabetes compared to participants without diabetes, and 0.93  for high vs. low physical activity. These results suggest a positive association between adiposity and diabetes and the risk of kidney stones, but no association with physical activity.We conducted a systematic review and meta-analysis to clarify the association between adiposity, diabetes, and physical activity and the risk of kidney stones. PubMed and Embase were searched up to April 22nd 2018 for relevant studies. Summary relative risks (RRs) and 95% confidence intervals (CIs) were calculated using random effects models. Thirteen cohort studies were included. The summary relative risk was 1.21  contains supplementary material, which is available to authorized users. Kidney stones are a major cause of morbidity and affects approximately 1–15% of the population around the world . The eco2) , pheterogeneity < 0.0001) . There was no indication of publication bias with Egger’s test, p = 0.28, or with Begg’s test, p = 0.17 . There was evidence of a nonlinear association between BMI and risk of kidney stones, pnonlinearity < 0.0001, and the curve was flat between 20 and 22, and increased at BMI values above this range  –16, 24 w2 = 0% [95% CI 0–79%], pheterogeneity = 0.59)  , 16, 17 59) Fig. A. The su2 = 67% [95% CI 0–90%], pheterogeneity < 0.0001)   were inc01) Fig. A. The su2 = 86% [95% CI 59–95%], pheterogeneity < 0.0001)   were inc2 = 51% [95% CI 0–76%], pheterogeneity = 0.03) .Ten cohort studies (seven publications) , 19–24 w2 = 80% [95% CI 46–92%], pheterogeneity < 0.0001)   , 27 wereheterogeneity = 0.003) with stronger associations among studies with self-reported data vs. those with measured data. There was also heterogeneity by adjustment for alcohol (pheterogeneity = 0.007), adjustment for diabetes (pheterogeneity = 0.002), thiazide use (pheterogeneity = 0.003), fluids (pheterogeneity = 0.003), sodium (pheterogeneity = 0.007), calcium supplements (pheterogeneity = 0.003), and physical activity (pheterogeneity = 0.003) with weaker associations among studies with adjustment for diabetes, calcium supplements and physical activity versus studies without such adjustment and stronger associations among studies with adjustment for alcohol, thiazide use, fluids, and sodium than among studies without such adjustment  having affected the findings, however, the association between BMI and kidney stones persisted in several subgroup analyses with adjustment for a number of confounding factors and there was no between subgroup heterogeneity. Nevertheless, confounding by risk factors that were not included in the statistical analyses or by risk factors that are not yet known is still possible. Measurement errors could have influenced the findings. When the analysis was stratified by whether the anthropometric factors were measured or self-reported the summary estimates were stronger among the studies with self-reported anthropometric measures compared to those with measured data, which might suggest overestimation of the association when using self-reported anthropometrics. However, it is also possible that this difference is due to differences in other study characteristics between studies. Ideally one should compare analyses of measured and self-reported data within the same study. As a meta-analysis of published literature it is possible that publication bias may have affected our findings, but we did not find evidence of such bias in the analysis of BMI or diabetes with the statistical tests used, however, the limited number of studies in the remaining analyses did not permit formal testing. There was high heterogeneity in the main analyses for BMI, weight and weight gain, however, this appeared to be explained to a large extent by one or two studies  in each Weight, weight gain and waist circumference are highly correlated with BMI and because none of the studies made further mutual adjustments between adiposity variables (for example between BMI and waist circumference) we are not able to conclude whether general or abdominal adiposity is more or equally important in increasing the risk of kidney stones and this is something that could be further clarified in future studies.Strengths of our analysis include the prospective design of the studies which therefore avoids recall bias, which can affect retrospective case–control studies, and limits the possibility that selection bias explains the results. The meta-analysis included > 10,300 cases among > 458,000 participants in the BMI analysis and > 13,400 cases and > 666,000 participants in the diabetes analysis, providing sufficient statistical power to detect even modest associations. To increase comparability between studies we conducted linear and nonlinear dose–response analyses. The results persisted in a number of subgroup and sensitivity analyses, suggesting that the findings are not likely to be completely explained by confounding. The results were robust to the influence of single studies, and the included studies had a relatively high study quality.Several potential mechanisms could explain an association between body fatness and increased risk of kidney stones. Obesity has been associated with high serum uric acid and gout , 46, galThe current findings have important public health implications in light of the current epidemics of overweight and obesity  and diabOur findings confirm that body fatness is an important risk factor for kidney stones and there was a 1.4-fold increase in risk among overweight and 2–3 fold increase in risk among obese and severely obese participants, respectively, while the lowest risk observed in the BMI range between 20 and 22. In addition, both waist circumference, weight and weight gain was associated with an increased risk, although additional studies are needed because of the limited data on these adiposity measures. A diagnosis of diabetes was associated with a 16% increase in the relative risk compared to persons without diabetes. No association was observed between physical activity and kidney stones, however, the possibility of a nonlinear association with reductions at low-to-moderate levels of activity cannot be excluded, but further studies are needed.Contribution: DA conceived of and designed the study. DA conducted the literature search and analyses and wrote the first draft of the paper. DA and MSY conducted the literature screening. All authors interpreted the data, contributed to the draft of the paper, revised the subsequent drafts for important intellectual content, read and approved the final manuscript. D. Aune takes primary responsibility for the integrity of the data and the accuracy of the data analysis.Supplementary material 1 (PDF 340 kb)Below is the link to the electronic supplementary material.


Polymorphisms have been identified to predispose to NAFLD. Here, we accessed the seven polymorphisms of rs1260326, rs780094 in GCKR, rs2954021 near TRIB1, rs2228603 in NCAN, rs58542926 in TM6SF2, rs12137855 near LYPLAL1, and rs10883437 near CPN1 on NAFLD susceptibility in the Uygur population.We collected 620 samples (317 NAFLD and 303 controls) for this case-control study. Meta-analysis was performed using Stata Software.Our data detected that the rs1260326  and rs780094  were significantly associated with the susceptibility to NAFLD in Uygur population. The rs1260326 and rs780094 T/T genotype are significantly associated with soda, egg, and soybean intakes in the consumption group with twice or more in a week. Furthermore, a significant haplotype effect of rs1260326/T- rs780094/T was found  compared with CC haplotype. An additional meta-analysis using 4352 cases and 10,168 controls established that rs780094  is significantly associated with NAFLD. Finally, among the 4 case-control studies on rs1260326, including 712 NAFLD and 658 controls, significant associations were found in Asian, liver biopsy, adult and pediatric groups.Collectively, both our case-control study and meta-analysis confirm a significant association between rs780094 and NAFLD. Additionally, our results suggest Asian-specific, liver biopsy-specific, adult-specific and pediatric-specific associations between the rs1260326 and NAFLD. Moreover, the rs1260326 and rs780094 T/T genotype are significantly associated with food habits, such as soda, egg, and soybean.The online version of this article (10.1186/s12944-018-0877-3) contains supplementary material, which is available to authorized users. Non-alcoholic fatty liver disease (NAFLD), one of the most common forms of chronic liver diseases, is the epidemic hepatic manifestation of the metabolic syndrome with hyperglycemia, dyslipidemia, and subclinical inflammation , 2. It mIn addition to environmental factors, recent studies suggest that genetic factors are involved in the development and progression of NAFLD , 7. The Therefore, in this study, we focus on the seven polymorphisms of rs1260326 (p.Leu446Pro), rs780094 in GCKR, rs2954021 near TRIB1, rs2228603 (p.Pro92Ser) in NCAN, rs58542926 (p.Glu167Lys) in TM6SF2, rs12137855 near LYPLAL1, and rs10883437 near CPN1, and assess the associations between these gene polymorphisms and NAFLD risks in the Chinese Uygur population.Written informed consent was obtained from all participants. All participants explicitly provided permission for DNA analyses as well as collection of relevant clinical data. This study was approved by the Ethics Committee of Xinjiang Medical University, Urumqi, China. It was conducted according to the standards of the Declaration of Helsinki.Subjects were from a Uygur population who lived in the Xinjiang Uygur Autonomous Region of China. We recruited the NAFLD group from Xinjiang Medical University between since January 2015 and January 2016, and the control group came from the same hospital in the same period. In the NAFLD group, there were 317 Uygur patients, mean age (42.92 ± 9.44) years with features of NAFLD and ultrasonographic (US) examinations. Inclusion criteria were: diagnosed in accordance with the standards set described previously [All patients completed the standard test registration form, and disclosed the following data: (1) General information: age, sex, race, Body mass index (BMI). (2) Special test: serum triglyceride (TG), total cholesterol (TC), high-density lipoprotein (HDL), low-density lipoprotein (LDL), serum uric acid (SUA), fasting plasma glucose (FPG), aspartate aminotransferase (AST), alanine aminotransferase (ALT), blood urea nitrogen (BUN), serum creatinine (SCr), Adiponectin (ADP), Retinol-binding protein 4 (RBP4), Cytokeratin 18-M30 (CK18-M30), and Cytokeratin 18-M65 (CK18-M65). The ADP, RBP4, CK18-M30 and CK18-M65 concentrations in serum were determined by ELISA Kit  according to the manufacture’s instruction.2O. This mixture was placed in the thermal cycler and underwent 25 cycles of 96 °C for 10 s, 50 °C for 5 s, and 60 °C for 30 s. When completed, 0.5-U SAP was added and the reaction mixture was incubated for 60 min. Prior to loading onto the PRISM 310, 10 ul of formamide was added to 1 ul of reaction mixture and samples were heated to 95 °C for 5 min. Finally, the primary data was analyzed by GeneMapper 4.0 .Genomic DNA was extracted from whole-blood samples using standard procedures (Promega). We used polymerase chain reaction (PCR)–ligase detection reaction (LDR) method to genotype the six polymorphisms. PCR-LDR reactions were performed as described by the manufacturer  , 19, witAn independent systematic literature search of studies in all languages to October 2017 across PubMed, EMBASE, Medline, Web of Science, Springer, Cochrane Library, ScienceDirect, and China National Knowledge Infrastructure (CNKI) was conducted. The search terms used were“(GCKR or glucokinase regulatory) and (fatty liver or NAFLD) and (genetic variants or genetic variations or SNPs)”. References from eligible literature were scanned to avoid missing studies. The details on the inclusion criteria included as follow: 1) only the case-control studies on the association between GCKR (rs1260326 and rs780094) and NAFLD were included; 2) the eligible studies must contain enough information for the calculation of odds ratio; 3) the trial should be included the underlying NAFLD as the outcome of study. Two authors (Wen CAI and Ping YAN) independently extracted data included the first author’s name, publication year, country, ethnic group, number of alleles or genotypes and the total number of cases and controls. A total of 44 literatures were identified initially from the search strategy above. After screening titles and abstracts, 34 full texts were then retrieved for details review. At last, the remaining 17 case-control studies were qualified for our meta-analysis  are presented as means ± standard deviation (S.D.). The difference between the NAFLD and control groups was analyzed using a Student’s t-test, Pearson’s Chi-squared test or the nonparametric Mann–Whitney U tests, as appropriate. The potential relationship of genotypic frequencies of the polymorphisms with the risk of NAFLD was evaluated by the odds ratios (ORs) with their 95% confidence intervals (CIs) from logistic regression models. Meta-analysis was performed using the Stata software  set to the fixed-effect or random-effect method according to the heterogeneity. Sensitivity analyses were performed by excluding one study at a time to evaluate the influence of single studies on summary HRs. To evaluate the publication bias, Begg’s funnel plot was assessed . All staP < 0.05).A total of 620 subjects were enrolled, consisted of 317 NAFLD and 303 healthy controls in this case–control study. Table The results for the observed and expected values of the genotypes at the polymorphisms locus rs1260326, rs780094, rs2954021, rs2228603, rs58542926, rs12137855, and rs10883437 in the control group were in H-W equilibrium, indicating that the samples in these groups were representative of the population, as shown in Table p = 0.03), the recessive model , and the variant allele vs. wide-type allele , showed a significant difference between NAFLD and control participants. In addition, for rs780094, the variant homozygote vs. wide-type homozygote , the dominant model , and the variant allele vs. wide-type allele , showed a significant difference between NAFLD and control participants. However, logistic regression analyses revealed that the five polymorphisms  were not associated with the risk of NAFLD  and soybean  intakes demonstrated increased risk of NAFLD; the rs780094 T/T genotype group among the soda , egg  and soybean  intakes also demonstrated increased risk of NAFLD.Table To evaluate the correlations of the SNPs in GCKR, we performed haplotype analysis between NAFLD and healthy controls. There are total of four common haplotypes (> 1%) among controls. The haplotype CC, compared with the other three haplotypes, was demonstrated more frequently in both NAFLD and healthy controls. And using the most common haplotype as reference, a significant haplotype effect of rs1260326/T- rs780094/T was found  (Table P > 0.05)  and each genotype, we calculated the clinical characteristics for each genotype. The genotypes of polymorphism rs1260326 include three genotypes of CC/CT/TT, so is rs780094. However, the differences of various clinical indicators among different groups were not statistically significant . And subgroup meta-analysis by age has lowered the heterogeneity . Furthermore, the adult subgroup and pediatric group showed significant association . Similarly, stratified analysis by ethnicity, liver disease, and NAFLD assessment showed significant associations existed among Asian group , NAFLD group  and liver biopsy group  . Furthermore, sensitivity analysis also revealed that the conclusion was not biased by any individual study  [GCKR, a negative regulator of glucokinase, regulates phosphorylation of glucose, glycolysis, and fatty acid synthesis in the liver , 36. A n = 0.82) , 39, 40.However, inconsistent with the previous findings that GCKR rs1260326 and rs780094 were significantly associated with insulin, triglyceride and fasting plasma glucose levels , we did Additionally, our study revealed that increased risk estimates were observed for interaction of rs1260326 T/T genotype with egg and soybean and rs780094 T/T genotype with soda, egg and soybean in the consumption group with twice or more in a week. Thus, understanding the mechanism of rs1260326 T/T and rs780094 T/T genotypes with soda, egg and soybean interaction will require further studies. Moreover, the correlation among ADP, RBP4, CK18, rs1260326, rs780094 and NAFLD was investigated in this study. The results showed a negative correlation between ADP and NAFLD; a positive correlation between RBP4 and NAFLD; a positive correlation between CK18-M30 and NAFLD. However, the differences of ADP, RBP4, CK18-M30 and CK18-M65 among different genotypes of rs1260326 and rs780094 in NAFLD patients were not statistically significant, which suggested the ADP, RBP4 and CK18-M30 might not be functioned by rs1260326 and rs780094 in the formation of NAFLD, and further fine-mapping studies in the susceptible region of the variants and more NAFLD-related serum cytokines are needed.2 = 75.1%, P = 0.001). Of note, when stratified by age, the heterogeneity was lower among the two subgroups with adverse significant associations, suggesting age difference may exist on GCKR rs1260326. Additionally, significantly increased risk of NAFLD among Asian population and liver biopsy groups were found in GCKR rs1260326. However, the limited sample size may have underestimated subtle effects of the genetic variant, and the above conclusions should be carefully considered.In the present study, we further do the meta-analysis for the two significant polymorphisms (rs1260326 and rs780094). The current meta-analysis includes 17 studies comprised of 4603 cases and 10,325 controls. The pooled results revealed that the GCKR rs780094 polymorphism was associated with increased risk of NAFLD. Moreover, stratified analysis by ethnicity, liver disease, source of control, NAFLD assessment, age, and obesity has demonstrated all subgroups were significantly correlated with increased risk of NAFLD, suggesting GCKR rs780094 may have a high effect on NAFLD incidence. Unexpectedly, we found no significant associations between GCKR rs1260326 and risk of NAFLD for the overall results with significant heterogeneity  polymorphism at allele model (T vs. C). (JPEG 2020 kb)