Skip to main content
NCBI home page
Wiley Open Access Collection logoLink to Wiley Open Access Collection
. 2016 Aug 11;106(9):773–788. doi: 10.1002/bdra.23540

Association between IRF6 and 8q24 polymorphisms and nonsyndromic cleft lip with or without cleft palate: Systematic review and meta‐analysis

Kachin Wattanawong 1,2, Sasivimol Rattanasiri 1,✉, Mark McEvoy 3, John Attia 3, Ammarin Thakkinstian 1
PMCID: PMC5095821  PMID: 27511269

Abstract

Background

We conducted a systematic review and meta‐analysis of interferon regulatory factor 6 and 8q24 polymorphisms with nonsyndromic cleft lip with/without cleft palate (NSCL/P).

Methods

Data extraction was independently performed by two reviewers. Genotypic effects of four polymorphisms from 31 studies were pooled separately by ethnicity using a mixed‐effect logit model with accounting for heterogeneity.

Results

For rs2235371, AA and GA carried, respectively, 51% (95% confidence interval [CI], 37%–61%) and 42% (95% CI, 32%–50%) lower risks of NSCL/P than GG genotypes in Asians, but these genotypes were not significant in Caucasians. For rs2013162, only AA was significant, that is, carried 0.65 (95% CI, 0.52–0.82) times lower odds than CC in Caucasians but not for Asians. For rs642961, AA and GA genotypes, respectively, carried 2.47 (95% CI, 1.41–4.35) and 1.40 (95% CI, 1.12–1.75) times higher odds in Asian, and 2.03 (95% CI, 1.52–2.71) and 1.58 (95% CI, 1.37–1.82) times higher odds in Caucasians compare with GG genotypes. For rs987525, AA and CA genotypes carried 2.27 (95% CI, 1.43–3.60) and 1.34 (95% CI, 1.02–1.77) times higher odds in Asian, and 5.25 (95% CI, 3.98–6.91) and 2.13 (95% CI–1.82, 2.49) times higher odds in Caucasians, and 1.42 (95% CI, 1.10–1.82) and 1.28 (95% CI, 1.09–1.50) times higher odds in mixed ethnicities compared with CC genotypes. These variant effects remained significant based on applying Bonferroni corrected‐thresholds, except in the mixed ethnicity.

Conclusion

We show robust variant effects in NSCL/P. Considering them with other genes and risk factors might be useful to improve prediction of NSCL/P occurrence. Birth Defects Research (Part A) 106:773–788, 2016. © 2016 The Authors Birth Defects Research Part A: Clinical and Molecular Teratology Published by Wiley Periodicals, Inc.

Keywords: IRF6, 8q24, nonsyndromic cleft lip with or without cleft palate, meta‐analysis, polymorphism

Introduction

Nonsyndromic cleft lip with or without cleft palate (NSCL/P) is one of the commonest birth malformations and varies by geographic origin and ethnicity (Bender, 2000). The incidence is approximately 1 in 500 live births in Asia and America, 1 in 1000 in Europe, and 1 in 2500 in Africa. Cleft lip without cleft palate is more frequent in males than females, with a ratio of 2:1, and frequently occurs on the left rather than right side (Mossey et al., 2009).

Patients with NSCL/P suffer from functional and cosmetic problems, and NSCL/P can affect speech and communication, which results in delayed development, low self‐confidence, and poor quality of life for both patient and family members. Although reconstruction with maxillo‐facial surgery is possible, severe cases sometimes require multiple corrective surgeries (e.g., cheiloplasy, palatoplasty, rhinoplasty, orthognathicsurgery), dental work, and speech rehabilitation. Thus, this condition places a significant burden on family and medical services (Strauss, 1999; Murray, 2002).

There have been attempts to identify disease genes associated with NSCL/P as identification of disease genes may shed light on the etiology of the condition and facilitate efforts at prevention of disease. Mutations of interferon regulatory factor 6 (IRF6), located on chromosome 1q32.2 (OMIM # 607199), is associated with the autosomal‐dominant Van der Woude syndrome, the most common Mendelian syndrome that has the cardinal signs of cleft lip with or without cleft palate (CL/P) and/or cleft palate only (CPO) with dental anomalies and pitted lips (Kondo et al., 2002; Rizos and Spyropoulos, 2004). Polymorphisms in IRF6 were later identified by Zucchero et al. (2004) and replicated by individual studies (Zucchero et al., 2004; Rahimov et al., 2008; Birnbaum et al., 2009b) and genome‐wide association studies (Birnbaum et al., 2009b; Grant et al., 2009; Beaty et al., 2010; Ludwig et al., 2012), as associated with NSCL/P. Animal studies also indicate that IRF6 is involved in proliferation‐differentiation of the keratinocyte (Richardson et al., 2009), and hyper‐proliferation of the epidermis may result in failure of terminal differentiation and multiple epithelial adhesion leading to CL/P (Ingraham et al., 2006; Richardson et al., 2006).

A genome‐wide association study of German subjects of Central European origin identified 146 polymorphisms in the 8q24 region, of which rs987525 (OMIM # 612858) was the most significantly associated with NSCL/P and was the major susceptibility locus in the general population (Birnbaum et al., 2009b). Grant et al. (2009) also conducted a case–control genome‐wide association studies (GWAS) in the United States and confirmed the associations of this region. In addition, Beaty et al. (2010) had replicated the GWAS in Asian subjects and had similar results, that is rs987525 was the most significant polymorphism. However, the biological mechanism of this SNP remains unclear.

Several independent studies replicated the effects of the IRF6 gene (at rs2235371 and rs642961) and the 8q24 region (at rs987525) on NSCL/P, and their effects were later pooled by a meta‐analysis (Wang et al., 2012). However, this meta‐analysis pooled only allele effects rather than genotypic effects, which assumes an additive mode of action. In addition, further studies have been published since then. We, therefore, performed an updated meta‐analysis on the same polymorphisms with additional IRF6 polymorphisms at rs2013162 aiming to assess both variant effects and possible genetic models.

Materials and Methods

Search Strategy

This review was conducted following the HuGE Review guideline (Little and Higgins, 2006; Gwinn et al., 2014). Studies were identified by two reviewers (K.W. and S.R.) using Medline (by means of PubMed) and Scopus (Sci Verse Scopus; Elsevier B.V.) databases up to February 15, 2016. Search terms used were as follows: (irf6 or irf‐6 or “interferon regulatory factor 6” or “interferon regulation factor‐6” or 8q24) and (polymorphism or gene or allele or SNP) and (NSCL/P or “cleft lip” or CL or “cleft palate” or CP or CPO or “nonsyndromic cleft”). Details of the search strategies are described in Supplementary Table 1, which is available online.

Inclusion and Exclusion Criteria

Two reviewers (K.W. and S.R.) independently selected studies by looking through all titles and abstracts of the identified studies. Any human population‐based (not familial‐based) association study published in English was included if it met the following criteria: the outcome of interest was NSCL/P, the studied polymorphisms were in IRF6 or 8q24, and sufficient data were reported, that is, allele/genotype frequency between case and control; or odds ratio (OR) of allele/genotype effects on outcome along with its 95% confidence interval (CI). The studies with insufficient information were excluded if authors did not provide data after two contacts. Where there were multiple publications with the same or overlapping subjects (which could be detected by incorporating information on authorship lists, setting, study period, genotyping method, and subjects) the most complete and/or recent results were used. The reference lists of the selected articles were also reviewed to identify additional relevant publications.

Data Extraction

Summary data for IRF6 and 8q24 were extracted independently by two reviewers (K.W. and S.R.) using a standardized data extraction form. Data on co‐variables, for example, mean age, percentage of males, ethnicity, and family history were also extracted. The main data (generic data and risk of bias assessment) were computerized separately by reviewer. Then, the two data were compared and validated to identify disagreement between. Meetings between the two reviewers (K.W. and S.R.) and the third party (A.T.) were then setup to discuss and solve disagreement.

Risk of Bias Assessment

The quality of studies was independently assessed by 2 reviewers (K.W. and S.R.) using a risk of bias assessment for genetic association studies, described in detail previously (Thakkinstian et al., 2005b, 2011). Briefly, the assessment considered five domains: selection bias, information bias, population stratification, selective reports, and assessment of Hardy‐Weinberg equilibrium (HWE). Each item was classified as yes, no, or unclear, corresponding to low risk of bias, high risk of bias, or unclear/insufficient information. Disagreement was solved by discussion with senior author (A.T.).

Outcome and Studied Gene of Interests

Our review focused on NSCL/P as the outcomes of interest, which diagnosed according to original studied. The studied genes were three polymorphisms within the IRF6 gene (rs2235371 G>A, rs642961 G>A, and rs2013162 C>A), and one located within 8q24 (rs987525 C>A).

Statistical Analysis

Data analysis was performed following a standard method of meta‐analysis for genetic association studies (Thakkinstian et al., 2005a, 2012). HWE was checked in control groups for case‐control studies using the Fisher's exact test. Data were then pooled based on studies that complied with HWE and if there were at least three studies for each studied polymorphism. A prevalence of minor allele frequency (MAF) for each polymorphism was estimated for each study, and then pooled across studies separately by ethnicity (Thakkinstian et al., 2005b).

Variant effects on NSCL/P were assessed using a per‐allele and per‐genotype approach. For the per‐allele approach, an OR of a versus A, if a and A were, respectively, minor and major alleles, along with its variance were estimated. Heterogeneity was then assessed using Cochrane's Q test and I2 statistic. The ORs were then pooled using a random‐effect model by Der Simonian & Laird method if heterogeneity was present (p value < 0.10 or I2 ≥ 25%); otherwise a fixed‐effect model with inverse‐variance method was applied.

For the per‐genotype approach, OR1 (aa vs. AA) and OR2 (Aa vs. AA) were estimated for each polymorphism across included studies. Heterogeneity was then assessed using the same methods described above. If any of the two ORs was heterogeneous, data were then pooled using a mixed‐effect hierarchical logit model (Thakkinstian et al., 2012). In this model, the genotypes were considered as fixed effects, whereas the study was considered as a random effect. A likelihood ratio test was used to assess whether an overall variant effect was significant. The pooled ORs along with 95% CI were then estimated by exponential coefficients of the mixed‐effect logit model. If the overall variant effect was present, the mode of genetic effect was captured by the parameter lambda (λ = logOR2 /logOR1), which was then estimated using the model‐free Bayesian approach (Minelli et al., 2005). The estimated λ (range: 0 to 1) would suggests a recessive, dominant, and additive effect if λ closes to 0, 1, and 0.5, respectively.

A sensitivity analysis was performed by including studies not observing HWE in the main pooling. Publication bias was assessed using the Egger's test and funnel plot (Egger et al., 1997). In addition, contour enhanced‐funnel plots were generated if there was any evidence of asymmetry suggested by funnel plot or Egger's test (Peters et al., 2008).

Analyses were performed using STATA version 14 (StataCorp, 2013; van Rooij et al., 2004) and WinBugs 1.4.2 (Spiegelhalter et al., 2007) with normal vague prior distributions for estimation of lambda and ORs. The model was run with 1000 iteration burn in, following by 10,000 iterations for estimation of parameters. A p‐value less than 0.05 was considered statistically significant, except for tests of heterogeneity where a level of 0.10 was used. In addition, Bonferroni correction was applied to adjust for 27 multiple tests in total (see Supplementary Table 2), which resulted in a Bonferroni corrected threshold of 1.85 × 10‐3.

Results

Identifying Studies

A total of 307 and 1155 studies were, respectively, located from Medline by means of PubMed and Scopus (Fig. 1). After 282 duplicates were removed, 1180 studies were screened on titles or abstracts, and 34 studies were eligible. Reasons for ineligibility are provided in Figure 1, but were mainly nonrelevant genes or populations, and nongenetic association studies. For studies meeting inclusion criteria, the study by Letra et al. (2012) had insufficient data but the author provided additional data after our request. Studied polymorphisms and number of studies for each polymorphism are described in Figure 1. Among them, three polymorphisms (i.e., rs2235371, rs2013162, rs642961) for IRF6, and one polymorphism (i.e., rs987525) for 8q24 had sufficient numbers of studies; therefore, further pooling was focused on only these polymorphisms from 31 studies.

Figure 1.

Figure 1

Open in a new tab

Flow chart for searching.

The characteristics of these 31 studies (Srichomthong et al., 2005; Rahimov et al., 2008; Jugessur et al., 2008; Ali et al., 2009; Birnbaum et al., 2009a, 2009b; Grant et al., 2009; Huang et al., 2009; Nikopensius et al., 2009; Tang et al., 2009; Carter et al., 2010; Mostowska et al., 2010; Pan et al., 2010; Paranaiba et al., 2010; Rojas‐Martinez et al., 2010; Shi et al., 2011; Weatherley‐White et al., 2011; Brito et al., 2012a, 2012b; Letra et al., 2012; Xu et al., 2012; Velazquez‐Aragon et al., 2012; Hikida et al., 2012; Bagordakis et al., 2013; Song et al., 2013; Lu et al., 2013; Aldhorae et al., 2014; Krasone et al., 2014; do Rego Borges et al., 2015; Kerameddin et al., 2015; Mijiti et al., 2015) are given in Table 1. All studies were case–control designs. Thirteen, 11, and 7 studies were conducted in Asian, Caucasian, and mixed populations. Mean age ranged from approximately 2.6 to 17.3 years, and percent male ranged from 46% to 74%. Fifteen, 9, 11, and 12 studies reported data on rs2235371, rs2013162, rs642961 for IRF6, and rs987525 for 8q24, with the outcome NSCL/P, respectively.

Table 1.

Characteristics of Included Studies

Author Year Country Ethnicity Study design SNP Locus Percentage male Percentage CLO
Srichomthong C. 2005 Thai Asian Case‐control rs2235371 IRF6 0.58 0.38
Jugessur A. 2008 Norway Caucasian Case‐control rs2235371 IRF6
rs2013162 IRF6
Rahimov F. 2008 Norway, Denmark Caucasian Case‐control rs642961 IRF6 0.4
Ali A. 2009 India Asian Case‐control rs2235371 IRF6 0.7
Birnbaum S. 2009 Central Europe Caucasian Case‐control rs2235371 IRF6
rs2013162 IRF6
rs642961 IRF6
Birnbaum S. 2009 Central Europe Caucasian Case‐control rs987525 8q24 0.22
Grant S. 2009 USA Caucasian Case‐control rs987525 8q24 0.69
Huang Y. 2009 China Asian Case‐control rs2235371 IRF6 0.15
rs2013162 IRF6
Nikopensius T. 2009 Estonia, Lithonia Caucasian Case‐control rs987525 8q24
Tang W. 2009 China Asian Case‐control rs2235371 IRF6 0.74
Carter TC. 2010 Ireland Caucasian Case‐control rs2235371 IRF6
rs2013162 IRF6
Mostowska A. 2010 Poland Caucasian Case‐control rs642961 IRF6 0.51 0.11
rs987525 8q24
Pan Y. 2010 China Asian Case‐control rs2235371 IRF6 0.64 0.4
rs642961 IRF6
Paranaiba L. 2010 Brazil Caucasian Case‐control rs2235371 IRF6
rs642961 IRF6
Rojas‐Martinez A. 2010 Mexico Mixed Case‐control rs987525 8q24
Author Year Country Ethnicity Study design SNP Locus Mean age Percentage male Percentage CLO
Shi J. 2011 China Asian Case‐control rs2235371 IRF6 1 to 28 0.46 0.29
rs642961 IRF6
Weatherley‐White R. 2011 Kenya Mixed Case‐control rs2013162 IRF6
rs987525 8q24
Brito L. 2012 Brazil Mixed Case‐control rs642961 IRF6 0.19
Brito L. 2012 Brazil Mixed Case‐control rs987525 8q24
Hikida M. 2012 Japan Asian Case‐control rs987525 8q24
Letra A. 2012 Brazil Caucasian Case‐control rs2235371 IRF6 17.32 0.6
rs2013162 IRF6
Velazquez‐Aragon J. 2012 Mexico Mixed Case‐control rs2235371 IRF6
rs987525 8q24
Xu M. 2012 China Asian Case‐control rs987525 8q24 7.95 0.68 0.2
Bagordakis E. 2013 Brazil Mixed Case‐control rs2013162 IRF6 0.35
Lu Y. 2013 China Asian Case‐control rs2235371 IRF6
rs2013162 IRF6
Song T. 2013 China Asian Case‐control rs2235371 IRF6
Aldhorae KA. 2014 Arab Asian Case‐control rs642961 IRF6
rs987525 8q24
Krasone K. 2014 Latvia Caucasian Case‐control rs642961 IRF6
Do Rego Borges A. 2015 Brazil Mixed Case‐control rs642961 IRF6
rs987525 8q24
Kerameddin S. 2015 Iran Asian Case‐control rs642961 IRF6
Mijiti A. 2015 China Asian Case‐control rs2235371 IRF6
rs2013162 IRF6
Open in a new tab

Risk of Bias Assessment

The results of bias assessment are presented in Supplementary Table 3. All studies had low risks of bias related to ascertainment of NSCL/P and population stratification. Twenty‐nine of 31 studies assessed HWE. Most studies (97%) had low risk of selective reporting of outcomes. Twenty‐seven (87%) and 24 (77%) studies described control ascertainment and quality control for genotyping, respectively.

rs2235371

Fifteen studies (i.e., 9 Asians, 5 Caucasians, and 1 mixed ethnicity) assessed effect of rs2235371 polymorphism on NSCL/P. Allele frequencies in the NSCL/P and control groups are described, and genotype frequencies in the control groups complied with HWE for all studies (Supplementary Table 4). The MAF was pooled by ethnicity. This suggested the pooled prevalence of MAF A in Asians was 0.34 (95% CI, 0.28–0.40) in the control group and 0.26 (95% CI, 0.19–0.32) in NSCL/P groups, yielding a pooled OR of 0.66 (95% CI, 0.58–0.75, p = 1.27 × 10‐9), which was still significant based on Bonferroni corrected‐threshold. The MAF A in Caucasians was very rare and much lower than in Asians, that is, 0.03 (95% CI, 0.02–0.04) and 0.02 (95% CI, 0.01–0.04) in control and NSCL/P groups, respectively. This yielded a pooled OR of 0.69 (95%CI, 0.43–1.12; p = 0.131), but this was not statistically significant.

The genotype frequencies between NSCL/P (n = 3513) and control (n = 4756) groups along with ORs for all studies are described in Table 2. Because MAF was different between Asians and Caucasians, pooling genotype effect was, therefore, performed separately by ethnicity. Degrees of heterogeneity I2 for OR1 (AA vs. GG) and OR2 (GA vs. GG) were, respectively, 60.0% (Chi‐square = 19.98; d.f. = 8; p = 0.010) and 0% (Chi‐square = 3.70; d.f. = 8, and p = 0.883) for pooling 9 Asian studies (n = 1677 vs. 1717). The corresponding pooled ORs were 0.49 (95% CI, 0.39–0.63; p = 7.73 × 10‐9) and 0.58 (95% CI, 0.50–0.68; p = 1.50 × 10‐12), see Figure 2A,B. These genotype effects were still significant after applying Bonferroni corrected‐threshold, indicating that carrying AA and GA is associated with approximately 51% and 42% lower odds of having NSCL/P than carrying GG genotype. The estimated lambda was 0.71 (95% CI, 0.38–0.98), indicating that a mode of variant effect could be from additive to dominant effect.

Table 2.

Genotype Frequencies and Genotype Effects for IRF‐6 at SNPs 2235371 Between NSCL/P and Control Groups

CL/P Control
Year No. of subjects Genotype No. of subjects Genotype AA/GG GA/GG
Author GG GA AA GG GA AA OR1 95% CI OR2 95% CI
Asian
Srichomthong C. 2005 192 93 72 27 278 100 137 41 0.71 0.40, 1.24 0.57 0.38, 0.84
Ali A. 2009 323 252 65 6 214 141 66 7 0.48 0.16, 1.46 0.55 0.37, 0.82
Huang Y. 2009 257 133 106 18 174 57 79 38 0.20 0.11, 0.39 0.58 0.38, 0.88
Tang W. 2009 66 30 31 5 96 34 50 12 0.47 0.15, 1.50 0.70 0.36, 1.37
Pan Y. 2010 127 69 40 18 115 47 60 8 1.53 0.62, 3.81 0.45 0.26, 0.78
Shi J. 2011 173 82 69 22 154 63 71 20 0.85 0.42, 1.68 0.75 0.47, 1.19
Lu Y. 2013 236 122 96 18 400 152 180 68 0.33 0.19, 0.58 0.66 0.47, 0.94
Song T. 2013 203 114 72 17 226 93 106 27 0.51 0.26, 1.00 0.55 0.37, 0.83
Mijiti A. 2015 100 80 17 3 60 39 20 1 1.46 0.15, 14.52 0.41 0.20, 0.88
Pooled 0.49 0.39, 0.63 0.58 0.50, 0.68
Caucasian
Jugessur A. 2008 314 297 17 0 416 399 17 0 1.34 0.03, 67.87 1.34 0.68, 2.65
Birnbaum S. 2009 442 435 7 0 952 913 39 0 2.10 0.04, 105.89 0.40 0.18, 0.88
Carter TC. 2010 460 456 4 0 894 869 25 0 1.91 0.04, 96.15 0.34 0.12, 0.92
Paranaiba LM. 2010 177 159 18 0 126 113 13 0 0.71 0.01, 36.13 0.98 0.46, 2.05
Letra A. 2012 311 283 26 2 281 245 34 2 0.87 0.12, 6.19 0.66 0.39, 1.13
Pooled 0.75 0.27, 1.75 0.82 0.55, 1.16
Mixed ethnicity
Velazquez JA. 2012 132 75 50 7 370 163 162 45 0.34 0.15, 0.79 0.67 0.44, 1.02
Open in a new tab

Figure 2.

Figure 2

Open in a new tab

Forest plots for rs2235371 of IRF6.

Among 5 Caucasian studies (n = 1704 and 2669 for NSCL/P and control), heterogeneity was not present in OR1 but was present for OR2 with an I2 of 0% (Chi‐square = 0.29; d.f. = 4; p = 0.990) and 51.9% (Chi‐square = 8.32; d.f. = 4; p = 0.080), respectively. The pooled OR1 and OR2 were 0.75 (95% CI, 0.27–1.75; p = 0.506) and 0.82 (95% CI, 0.55–1.16; p = 0.262), see Figure 2C,D; indicating nonsignificant variant effects for both AA and GA genotypes when compared with GG genotypes.

Publication bias was assessed for both pooled ORs in Asian and Caucasian by funnel plot and Egger's test. There was no evidence of asymmetry of the funnel, see Supplementary Figure 1A–D. These results agreed with the Egger's tests, which indicated no evidence of asymmetry of funnels for both OR1 (coefficient = 1.448; p = 0.408) and OR2 (coefficient = ‐1.018; p = 0.358) in Asian, and OR1 (coefficient = 0.480; p = 0.261) and OR2 (coefficient = ‐2.763; p = 0.452) in Caucasian.

rs2013162

Among the nine studies that assessed the association between rs2013162 in IRF6 and NSCL/P, four studies, three studies, and two studies were conducted in Caucasians, Asians, and in mixed ethnicity, respectively. Allele effects were pooled in Caucasian and Asian where numbers were sufficient. All studies complied with HWE with the pooled MAF A of 0.35 (95% CI,0.34–0.37) and 0.42 (95% CI, 0.35–0.50), respectively, in Caucasian, Asian of control populations. The pooled ORs of A versus C alleles were, respectively, 0.84 (95% CI, 0.76–0.93; p = 4.16 × 10‐4) and 0.99 (95% CI, 0.57–1.73; p = 0.974) in Caucasians and Asians (Supplementary Table 5).

The genotype frequency between NSCL/P groups of 9 studies are described in Table 3. Among four Caucasian studies (n = 1530 vs. 2511 for NSCL/Ps vs. controls), ORs (i.e., OR1 [AA vs. CC] and OR2 [CA vs. CC]) showed low heterogeneity for OR1 (Chi‐square = 3.19; d.f. = 3; p = 0.363; I2 = 6.1%) and OR2 (Chi‐square = 0.57; d.f. = 3; p = 0.904; I2 = 0%), see Figure 3A,B. The pooled OR1 and OR2 were, respectively, 0.65 (95% CI, 0.52–0.82; p = 1.93 × 10‐4) and 0.89 (95% CI, 0.78–1.02; p = 0.102), suggesting that the odds of NSCL/P was significantly different in AA genotype even based on Bonferroni corrected threshold, but was not significantly different in CA genotype when compared with CC genotype. The estimated lambda was 0.32 (95% CI, 0.02–0.92), which possibly indicated between a recessive to additive effect of A allele.

Table 3.

Genotype Frequencies and Genotype Effects for IRF‐6 at SNPs 2013162 Between NSCL/P and Control Groups

CL/P Control
Year No. of subjects Genotype No. of subjects Genotype AA/CC CA/CC
Author CC CA AA CC CA AA OR1 95% CI OR2 95% CI
Asian
Huang Y. 2009 165 53 84 28 174 51 87 36 0.75 0.40, 1.40 0.93 0.57, 1.51
Lu Y. 2013 236 57 132 47 400 156 200 44 2.92 1.75, 4.87 1.81 1.24, 2.63
Mijiti A. 2015 100 44 41 15 59 16 31 12 0.46 0.18, 1,18 0.48 0.23, 1.01
Pooled 1.40 0.97, 2.01 1.20 0.92, 1.57
Caucasian
Jugessur A. 2008 313 136 141 36 416 174 190 52 0.89 0.55, 1.43 0.95 0.69, 1.30
Birnbaum S. 2009 460 215 210 35 952 410 430 112 0.60 0.39, 0.90 0.93 0.74, 1.18
Carter TC. 2010 447 204 202 41 861 350 409 102 0.69 0.46, 1.03 0.85 0.67, 1.08
Letra A. 2012 310 143 140 27 282 110 128 44 0.47 0.28, 0.81 0.84 0.60, 1.19
Pooled 0.65 0.52, 0.82 0.89 0.78, 1.02
Mixed ethnicity
Weatherley‐White RC. 2011 128 76 45 7 105 56 41 8 0.65 0.22, 1.88 0.81 0.47, 1.40
Bagordakis E. 2013 233 122 88 23 308 142 137 29 0.92 0.51, 1.68 0.75 0.52, 1.07
Open in a new tab

Figure 3.

Figure 3

Open in a new tab

Forest plots for rs2013162 of IRF6.

Among three Asian studies (n = 501 for NSCL/Ps and 633 for controls), heterogeneity was present in both OR1 and OR2 with an I2 of 88.4% (Chi‐square = 17.21; d.f. = 2; p < 0.001) and 82.6% (Chi‐square = 11.52; d.f. = 2; p = 0.003), respectively. The pooled OR1 and OR2 were 1.40 (95% CI, 0.97–2.01; p = 0.069) and 1.20 (95% CI, 0.92–1.57; p = 0.172), see Figure 3C,D; indicating nonsignificant variant effects for both AA and CA genotypes when compared with CC genotypes.

Publication bias was assessed for both pooled ORs in Caucasian and Asian by funnel plot and Egger's test. Funnel plots suggested little asymmetry (Supplementary Fig. 2A–D), but the Egger's test did not confirm this. The coefficients of asymmetry was ‐2.07 (SE = 5.25; p = 0.731) for OR1 and ‐0.25 (SE = 1.64; p = 0.894) for OR2 in Caucasian, and ‐8.58 (SE = 5.45; p = 0.360) for OR1 and ‐7.41 (SE = 1.88; p = 0.158) for OR2 in Asian.

rs642961

Eleven studies assessed effect of rs642961 polymorphism on NSCL/P. Among them, five studies, four studies, and two studies were conducted in Caucasians, Asians, and mixed ethnicity, respectively. One study (Brito et al., 2012a) did not comply with HWE and, thus, excluded from pooling. Data were sufficient for pooling in Caucasians and Asians. The pooled MAF A was 0.19 (95% CI, 0.17–0.22) and 0.17 (95% CI, 0.13–0.22) in Caucasians and Asians, respectively (Supplementary Table 6). The pooled ORs (A vs. G) were 1.50 (95% CI, 1.35–1.68, p = 5.11 × 10‐13) and 1.47 (95% CI, 1.09–1.98; p = 1.10 × 10‐2) in Caucasians and Asians, respectively.

The genotype frequencies between NSCL/P and control groups are described in Table 4. Genotype effects were sufficient for pooling in Caucasians and Asians. Among 5 Caucasians studies (n = 1244 vs. 3069 for NSCL/Ps and controls), the OR1 (AA vs. GG) and OR2 (GA vs. GG) were estimated with low heterogeneity for OR1 (Chi‐square = 5.31; d.f. = 4; p = 0.257; I2 = 24.6%) and moderate for OR2 (Chi‐square = 6.08; d.f. = 4; p = 0.194; I2 = 34.2%). The mixed‐effect logit model yielded pooled OR1 and OR2 of 2.03 (95% CI, 1.52–2.71; p = 1.81 × 10‐6) and 1.58 (95% CI, 1.37–1.82; p = 4.59 × 10‐10), which suggested that persons with AA and GA genotypes were 2.03 and 1.58 times significantly higher risk of NSCL/P than persons with GG genotype (Fig. 4A,B). These effects were still significant based on Bonferroni corrected threshold. The estimated lambda value was 0.62 (95% CI, 0.17–0.96) suggesting a potential mode of effect between recessive to additive effect.

Table 4.

Genotype Frequencies and Genotype Effects for IRF‐6 at SNPs 642961 Between NSCL/P and Control Groups

CL/P Control
Year No. of subjects Genotype No. of subjects Genotype AA/GG GA/GG
Author GG GA AA GG GA AA OR1 95% CI OR2 95% CI
Asian
Pan Y. 2010 127 67 54 6 115 81 33 1 7.25 0.85, 61.75 1.98 1.15, 3.40
Shi J. 2011 173 100 62 11 156 111 43 2 6.11 1.32, 28.21 1.60 1.00, 2.57
Aldhorae KA. 2014 242 159 74 9 420 311 101 8 2.20 0.83, 5.81 1.43 1.00, 2.05
Kerameddin S. 2015 150 88 51 11 150 84 57 9 1.17 0.46, 2.96 0.85 0.53, 1.38
Pooled 2.47 1.41, 4.35 1.40 1.12, 1.75
Caucasian
Rahimov F. 2008 368 180 155 33 1244 757 427 60 2.31 1.47, 3.65 1.53 1.19, 1.95
Birnbaum S. 2009 460 237 189 34 951 622 283 46 1.94 1.22, 3.10 1.75 1.38, 2.22
Mostowska A. 2010 165 82 74 9 565 349 184 32 1.20 0.55, 2.61 1.71 1.19, 2.46
Paranaiba LM. 2010 177 128 45 4 126 90 36 0 6.34 0.34, 119.19 0.88 0.53, 1.47
Krasone K. 2014 74 41 29 4 183 129 53 1 12.59 1.37, 115.80 1.72 0.97, 3.05
Pooled 2.03 1.52, 2.71 1.58 1.37, 1.82
Mixed ethnicity
Brito LA. 2012 471 320 123 28 391 285 90 16 1.56 0.83, 2.94 1.22 0.89, 1.67
do Rego Borges A. 2015 293 214 73 6 352 282 63 7 1.13 0.37, 3.41 1.53 1.04, 2.24
Open in a new tab

Figure 4.

Figure 4

Open in a new tab

Forest plots for rs642961 of IRF6.

Among four Asians studies (n = 300 vs. 271 for NSCL/Ps and controls), the OR1 (AA vs. GG) and OR2 (GA vs. GG) were estimated with moderate heterogeneity for OR1 (Chi‐square = 4.77; d.f. = 3; p = 0.189; I2 = 37.1%) and OR2 (Chi‐square = 5.93; d.f. = 3; p = 0.115; I2 = 49.4%). The mixed‐effect logit model yielded pooled OR1 and OR2 of 2.47 (95% CI, 1.41–4.35; p = 1.65 × 10‐3) and 1.40 (95% CI, 1.12–1.75; p = 3.02 × 10‐3), which suggested that persons with AA and GA genotypes were 2.47 and 1.40 times significantly higher risk of NSCL/P than persons with GG genotype (Fig. 4C,D). However, only OR1 was still significant based on Bonferroni corrected threshold. The estimated lambda value was 0.62 (95% CI, 0.17–0.96) suggesting a potential mode of variant effect could be from a recessive to additive mode.

No evidence of reporting bias was seen with either OR1 or OR2 suggested from Egger tests (in Caucasian: coefficient = 0.98; SE = 0.98; p = 0.391 for OR1 and coefficient = ‐1.61; SE = 1.68; p = 0.409 for OR2; in Asian: coefficient = 3.14; SE = 1.22; p = 0.124 for OR1 and coefficient = 0.66; SE = 5.33; p = 0.913 for OR2) and funnel plots (Supplementary Fig. 3A–D).

rs987525

Allele frequencies of 12 studies assessed effect of rs987525 polymorphism on NSCL/P and all observed HWE (Supplementary Table 7). The MAF A in controls were 0.15 (95% CI, 0.00–0.31) in three Asian studies, and 0.20 (95% CI, 0.17–0.22) in four Caucasian studies, and 0.27 (95% CI, 0.17–0.38) in five mixed ethnicity. The estimated ORs for allele A versus C in these corresponding ethnicities were 1.48 (95% CI, 1.21–1.81; p = 1.21 × 10‐4), and 2.20 (95% CI, 1.96–2.46; p < 1.00 × 10‐12), and 1.21 (95% CI, 1.09–1.36; p = 7.5 × 10‐4), respectively.

Among three Asian studies (n = 625 in NSCL/P and 811 in controls), the pooled OR1 (AA vs. CC) and OR2 (CA vs. CC) were, respectively, 2.27 (95% CI, 1.43–3.60; p = 4.71 × 10‐4) and 1.34 (95% CI, 1.02–1.77; p = 3.71 × 10‐2), with no heterogeneity for OR1 (Chi‐square = 0.10; d.f. = 2; p = 0.950; I2 = 0%) and OR2 (Chi‐square = 1.83; d.f. = 2; p = 0.401; I2 = 0%) (Table 5; Fig. 5A,B). This indicates that those with AA and CA genotypes were 2.27 and 1.34 times significantly, based on Bonferroni corrected threshold, more likely to have NSCL/P than those with the CC genotype. The estimated lambda was 0.45 (95% CI, 0.27–0.67), indicating an additive model was most likely.

Table 5.

Genotype Frequencies and Genotype Effects for 8q24 at SNPs 987525 Between NSCL/P and Control Groups

CL/P Control
Year No. of subjects Genotype No. of subjects Genotype AA/CC CA/CC
Author CC CA AA CC CA AA OR1 95% CI OR2 95% CI
Asian
Hikida M. 2012 167 145 21 1 190 170 20 0 3.52 0.14, 86.96 1.23 0.64, 2.36
Xu MY. 2012 216 188 26 2 200 175 24 1 1.86 0.17, 20.71 1.01 0.56, 1.82
Aldhorae KA. 2014 242 76 119 47 421 189 186 46 2.54 1.56, 4.13 1.59 1.12, 2.26
Pooled 2.27 1.43, 3.60 1.34 1.02, 1.77
Caucasian
Birnbaum S. 2009 462 172 228 62 952 604 312 36 6.05 3.88, 9.43 2.57 2.02, 3.26
Grant S. 2009 111 45 51 15 5951 3667 2009 275 4.45 2.45, 8.08 2.07 1.38, 3.10
Nikopensius T. 2009 217 114 82 21 1267 889 353 25 6.55 3.55, 12.08 1.81 1.33, 2.47
Mostowska A. 2010 165 77 70 18 565 357 186 22 3.79 1.94, 7.41 1.75 1.21, 2.52
Pooled 5.25 3.98, 6.91 2.13 1.82, 2.49
Mixed ethnicity
Rojas‐Martinez A. 2010 149 109 31 9 303 231 69 3 6.36 1.69, 23.95 0.95 0.59, 1.54
Weatherley‐White RC. 2011 129 42 63 24 105 38 50 17 1.28 0.60, 2.73 1.14 0.64, 2.03
Brito LA. 2012 667 278 302 87 589 284 239 66 1.35 0.94, 1.93 1.29 1.02, 1.64
Velazquez JA. 2012 132 89 42 1 370 267 91 12 0.25 0.03, 1.95 1.39 0.89, 2.14
do Rego Borges A. 2015 293 94 143 56 352 141 155 56 1.50 0.95, 2.36 1.38 0.98, 1.96
Pooled 1.42 1.10, 1.82 1.28 1.09, 1.50
Open in a new tab

Figure 5.

Figure 5

Open in a new tab

Forest plots for rs987525 of 8q24.

Among four Caucasian studies (n = 955 in NSCL/P and 8735 in controls), the pooled OR1 (AA vs. CC) and OR2 (CA vs. CC) were, respectively, 5.25 (95% CI, 3.98–6.91; p < 1.00 × 10‐12) and 2.13 (95% CI, 1.82–2.49; p < 1.00 × 10‐12), with no heterogeneity for OR1 (Chi‐square = 2.09; d.f. = 3; p = 0.553; I2 = 0%) and moderate heterogeneity for OR2 (Chi‐square = 4.50; d.f. = 3; p = 0.212; I2 = 33.3%) (Table 5; Fig. 5C,D). This indicates that those with AA and CA genotypes were 5.25 and 2.13 times significantly, based on Bonferroni corrected threshold, more likely to have NSCL/P than those with the CC genotype. The estimated lambda was 0.45 (95% CI, 0.27–0.67), indicating an additive model was most likely.

Among five studies with mixed ethnicity (n = 1370 in NSCL/P and 1719 in controls), the corresponding pooled ORs were 1.42 (95% CI, 1.10–1.82; p = 6.10 × 10‐3) and 1.28 (95% CI, 1.09–1.50; p = 2.47 × 10‐3) with moderate heterogeneity for OR1 (Chi‐square = 7.87; d.f. = 4; p = 0.096; I2 = 49.2%) and no evidence of heterogeneity for OR2 (Chi‐square = 1.92; d.f. = 4; p = 0.751; I2 = 0%) (Fig. 5F,G). These two ORs were not significant based on Bonferroni corrected threshold. The estimated lambda 0.68; 95% CI, 0–1), which could possibly be between additive to dominant mode.

There was no evidence of reporting bias (in Asian: Egger coefficient = ‐0.02; SE = 0.28; p = 0.951 for OR1 and coefficient = ‐2.50; SE = 1.29; p = 0.303 for OR2; in Caucasian: coefficient = ‐2.76; SE = 2.48; p = 0.380 for OR1 and coefficient = ‐3.93; SE = 2.47; p = 0.252 for OR2; in mixed ethnicity: coefficient = 0.10; SE = 1.45; p = 0.951 for OR1 and coefficient = ‐0.93; SE = 1.00; p = 0.422 for OR2; see Funnel plots in Supplementary Figure 4A–F.

Discussion

We conducted a systematic review and meta‐analysis to assess genetic effects of polymorphisms in IRF6 and 8q24 region on NSCL/P. A total of 9 to 15 studies were included in the pooling for three polymorphisms (i.e., rs2235371, rs2013162, rs642961) for IRF6 and for rs987525 in 8q24. The MAF for these corresponding polymorphisms were similar across ethnicities except for rs2235371 where the MAF was common in Asians but rare in Caucasians, whereas MAF for rs987525 was common in Caucasians but less common in Asians. Because the MAF of rs2235371 was different between ethnicities, the genotype effect for this SNP was different. All polymorphisms were significantly associated with NSCL/P in some ethnicities. A significant protective effect of rs2235371 in Asians and effect of rs2013162 in Caucasians were suggested, that is, carrying one copy of the A allele for both polymorphisms lowered the odds of NSCL/P compared with G and C alleles, respectively. In addition, our findings indicated risk effects of minor alleles for rs642961 and rs987525.

Our findings are consistent with previous human GWAS studies (Birnbaum et al., 2009b; Grant et al., 2009; Beaty et al., 2010; Ludwig et al., 2012). In addition, our findings are also consistent with the previous meta‐analysis of genetic association studies by Wang et al. (2012), that included 10, 6, and 8 studies in the pooling of rs2235371 and rs642961 for IRF6, and rs987525 for 8q24, respectively. We confirmed the previous findings by expanding to a larger number of included genetic association studies (i.e., 15, 11, and 12 studies for these corresponding polymorphisms) and applying Bonferroni correction for multiple tests. Although each polymorphism was considered separately in meta‐analysis for genetic association studies, this type of study had been used to replicate and confirm GWAS. Not only allele effects were estimated, but also pooled MAF and genotype effects by ethnicity additionally provided from previous evidences.

Formation of the lip and palate in human embryos starts at the fourth week and is complete by the twelfth week after fertilization. This process occurs by migration of the ectoderm, and formation of frontal, medial and lateral nasal processes, which then fuse together to form the normal lips and nose (Shkoukani et al., 2013). A normal palate also occurs after complete fusion of the primary and secondary palates; failure of fusion causes separation of the lip/s and/or palate, that is, cleft lip and cleft palate, respectively. Although the function of the IRF6 gene on the cleft lip formation is unclear, evidence has supported an effect of this gene on other birth malformations including Van der Woude syndrome and popliteal pterygium syndrome (Kondo et al., 2002; Kayano et al., 2003; Shotelersuk et al., 2003; Gatta et al., 2004; Ghassibé et al., 2004; Item et al., 2004). High IRF6 mRNA levels are also seen at the medial edge of the fusing palate (Kondo et al., 2002).

In addition, evidence from animal studies also supports the role of IRF6 on NSCL/P. Animal models have shown abnormalities of epithelial differentiation in mice with a mutation of IRF6 (Iwata et al., 2013). This supports a role for IRF6 in the formation and maintenance of the periderm, and spatiotemporal regulation for appropriate palatal adhesion (Richardson et al., 2009). Another study on mice with mutation of IRF6 showed abnormal skin, limb, and craniofacial development, which suggests a role for IRF6 in keratinocyte proliferation and differentiation of face and limb (Ingraham et al., 2006). The role of the 8q24 region on NSCL/P is still unclear but it might be involved in maintenance of an undifferentiated state in human acute myeloblastic leukemia cells (HL60) (Hirano et al., 2008), control of Myc expression (Uslu et al., 2014), or cranial facial enhancers (Attanasio et al., 2013).

The etiology of NSCL/P is still unclear but evidence suggests that both genetic and nongenetic factors play a role (Dixon et al., 2011; Shkoukani et al., 2013). Environmental factors may also play a role in NSCL/P. These factors may be independent or they may be modified by genes such as 5, 10‐methylenetetrahydrofolate reductase and folate (Bufalino et al., 2010). Evidence from cohort studies suggested that oral folate supplements before or during the periconception period might be able to reduce risk of NSCL/P (van Rooij et al., 2004; Kelly et al., 2012; Li et al., 2012); this is being confirmed by a randomized controlled trial (Wehby et al., 2012). Other factors associated with NSCL/P include alcohol consumption (Lieff et al., 1999; DeRoo et al., 2008), gestational diabetes (Correa et al., 2008), maternal age>40 years (Herkrath et al., 2012), and zinc deficiency (Warkany and Petering, 1972).

Our results also suggested that the IRF6 ((i.e., rs2235371, rs2013162, rs642961) and 8q24 (rs987525) SNPs contributed to NSCL/P, in which the rs987525 had highest contribution. This might be comparable to the effect of folate with an estimated PAR approximately 18% (van Rooij et al., 2004; Kelly et al., 2012). For public health prevention, screening these genes and other genes identified by previous meta‐analyses (e.g., VAX1, Figueiredo et al., 2014; or bone morphogenetic protein 4, Hu et al., 2015), and combining them with environmental factors (e.g., folate, alcohol, etc.) may lead to a useful risk score for prediction of NSCL/P occurrence (Seddon et al., 2009).

Our study has some strengths. We attempted to pool variant effects on NSCL/P separately by ethnicity if data were available. Prevalence of MAFs for each study polymorphism was pooled across studies and ethnicity. Mode of variant effect was estimated using a mixed‐effect logistic regression (Thakkinstian et al., 2005a) to capture lambda (Minelli et al., 2005). However, our study also has some limitations. First, given that we worked on summary data, we could not control for confounding effects, although the major source of confounding for genetic studies is population stratification and we summarized results by ethnic group. Second, we could only identify association between genes and NSCL/P, and we do not know whether the genes directly affect NSCL/P or whether their effects might be exerted through other environmental factors, such as folate. Exploring this would require individual patient data where data for folate and other co‐variables are also available. Third, we could not identify which polymorphism is the disease gene among those three polymorphisms in IRF6 gene because evidences showed linkage disequilibrium between rs2235371 and rs2013162 in Mexican (Ibarra‐Arce et al., 2015) and Chinese (Mijiti et al., 2015) population (r2 = 0.39 and 0.25, respectively) but not for rs642961 (Pan et al., 2010) (r2 = 0.07). To determine this required raw data for performing a haplotype analysis. Some variant effects were heterogeneous (i.e., varied) across studies; we could explore for a cause of heterogeneity by fitting percent male in a meta‐regression, but this was not detected (data were not shown). There might be other characteristics that cause heterogeneity, but those data were not available. Finally, there might be gene to gene interactions, but summary data do not allow us to explore this.

Conclusions

All polymorphisms, rs2235371, rs2013162, and rs642961 at IRF6 and rs987525 in 8q24 were significantly associated with NSCL/P. However, after applying Bonferroni corrected‐thresholds, variant effects remained significant, except in mixed ethnicity. We show robust variant effects in NSCL/P. Including them with other risk factors and other genes in a risk prediction model of NSCL/P might be useful to improve prediction of NSCL/P occurrence.

Supporting information

Additional Supporting information may be found in the online version of this article.

Supplement Figure 1. Funnel plots for rs2235371 of IRF6.

Click here for additional data file. (914.8KB, tif)

Supplement Figure 2. Funnel plots for rs2013162 of IRF6.

Click here for additional data file. (843.3KB, tif)

Supplement Figure 3. Funnel plots for rs642961 of IRF6.

Click here for additional data file. (1.5MB, tif)

Supplement Figure 4. Funnel plots for rs987525 of 8q24.

Click here for additional data file. (1.5MB, tif)

Supplement Tables.

Click here for additional data file. (58KB, docx)

Acknowledgments

This manuscript is part of Dr. Kachin Wattanawong's training for a Ph.D. (Clinical Epidemiology), the Faculty of Medicine Ramathibodi Hospital, Mahidol University, Bangkok Thailand. We thank Associate Professor Ariadne Letra, Department of Endodontics, School of Dentistry, the University of Texas Health Science Center at Houston for providing additional data.

References

  1. Aldhorae KA, Bohmer AC, Ludwig KU, et al. 2014. Nonsyndromic cleft lip with or without cleft palate in arab populations: genetic analysis of 15 risk loci in a novel case‐control sample recruited in Yemen. Birth Defects Res A Clin Mol Teratol 100:307–313. [DOI] [PubMed] [Google Scholar]
  2. Ali A, Singh SK, Raman R. 2009. MTHFR 677TT alone and IRF6 820GG together with MTHFR 677CT, but not MTHFR A1298C, are risks for nonsyndromic cleft lip with or without cleft palate in an Indian population. Genet Test Mol Biomarkers 13:355–360. [DOI] [PubMed] [Google Scholar]
  3. Attanasio C, Nord AS, Zhu Y, et al. 2013. Fine tuning of craniofacial morphology by distant‐acting developmental enhancers. Science 342:1241006. [DOI] [PMC free article] [PubMed] [Google Scholar]
  4. Bagordakis E, Paranaiba LM, Brito LA, et al. 2013. Polymorphisms at regions 1p22.1 (rs560426) and 8q24 (rs1530300) are risk markers for nonsyndromic cleft lip and/or palate in the Brazilian population. Am J Med Genet A 161:1177–1180. [DOI] [PubMed] [Google Scholar]
  5. Beaty TH, Murray JC, Marazita ML, et al. 2010. A genome‐wide association study of cleft lip with and without cleft palate identifies risk variants near MAFB and ABCA4. Nat Genet 42:525–529. [DOI] [PMC free article] [PubMed] [Google Scholar]
  6. Bender PL. 2000. Genetics of cleft lip and palate. J Pediatr Nurs 15:242–249. [DOI] [PubMed] [Google Scholar]
  7. Birnbaum S, Ludwig KU, Reutter H, et al. 2009a. IRF6 gene variants in Central European patients with non‐syndromic cleft lip with or without cleft palate. Eur J Oral Sci 117:766–769. [DOI] [PubMed] [Google Scholar]
  8. Birnbaum S, Ludwig KU, Reutter H, et al. 2009b. Key susceptibility locus for nonsyndromic cleft lip with or without cleft palate on chromosome 8q24. Nat Genet 41:473–477. [DOI] [PubMed] [Google Scholar]
  9. Brito LA, Bassi CF, Masotti C, et al. 2012a. IRF6 is a risk factor for nonsyndromic cleft lip in the Brazilian population. Am J Med Genet A 158A:2170–2175. [DOI] [PubMed] [Google Scholar]
  10. Brito LA, Paranaiba LM, Bassi CF, et al. 2012b. Region 8q24 is a susceptibility locus for nonsyndromic oral clefting in Brazil. Birth Defects Res A Clin Mol Teratol 94:464–468. [DOI] [PubMed] [Google Scholar]
  11. Bufalino A, Ribeiro Paranaiba LM, Nascimento de Aquino S, et al. 2010. Maternal polymorphisms in folic acid metabolic genes are associated with nonsyndromic cleft lip and/or palate in the Brazilian population. Birth Defects Res A Clin Mol Teratol 88:980–986. [DOI] [PubMed] [Google Scholar]
  12. Carter TC, Molloy AM, Pangilinan F, et al. 2010. Testing reported associations of genetic risk factors for oral clefts in a large Irish study population. Birth Defects Res A Clin Mol Teratol 88:84–93. [DOI] [PMC free article] [PubMed] [Google Scholar]
  13. Correa A, Gilboa SM, Besser LM, et al. 2008. Diabetes mellitus and birth defects. Am J Obstet Gynecol 199:237 e231–239. [DOI] [PMC free article] [PubMed] [Google Scholar]
  14. DeRoo LA, Wilcox AJ, Drevon CA, et al. 2008. First‐trimester maternal alcohol consumption and the risk of infant oral clefts in Norway: a population‐based case‐control study. Am J Epidemiol 168:638–646. [DOI] [PMC free article] [PubMed] [Google Scholar]
  15. Dixon MJ, Marazita ML, Beaty TH, et al. 2011. Cleft lip and palate: understanding genetic and environmental influences. Nat Rev Genet 12:167–178. [DOI] [PMC free article] [PubMed] [Google Scholar]
  16. do Rego Borges A, Sa J, Hoshi R, et al. 2015. Genetic risk factors for nonsyndromic cleft lip with or without cleft palate in a Brazilian population with high African ancestry. Am J Med Genet A 167a:2344–2349. [DOI] [PubMed] [Google Scholar]
  17. Egger M, Davey Smith G, Schneider M, et al. 1997. Bias in meta‐analysis detected by a simple, graphical test. BMJ 315:629–634. [DOI] [PMC free article] [PubMed] [Google Scholar]
  18. Figueiredo JC, Ly S, Raimondi H, et al. 2014. Genetic risk factors for orofacial clefts in Central Africans and Southeast Asians. Am J Med Genet A 164a:2572–2580. [DOI] [PubMed] [Google Scholar]
  19. Gatta V, Scarciolla O, Cupaioli M, et al. 2004. A novel mutation of the IRF6 gene in an Italian family with Van der Woude syndrome. Mutat Res 547:49–53. [DOI] [PubMed] [Google Scholar]
  20. Ghassibé M, Revencu N, Bayet B, et al. 2004. Six families with van der Woude and/or popliteal pterygium syndrome: all with a mutation in the IRF6 gene. J Med Genet 41:e15. [DOI] [PMC free article] [PubMed] [Google Scholar]
  21. Grant SF, Wang K, Zhang H, et al. 2009. A genome‐wide association study identifies a locus for nonsyndromic cleft lip with or without cleft palate on 8q24. J Pediatr 155:909–913. [DOI] [PubMed] [Google Scholar]
  22. Gwinn M, Ioannidis JP, Little J, et al. 2014. Editorial: updated guidance on human genome epidemiology (HuGE) reviews and meta‐analyses of genetic associations. Am J Epidemiol 180:559–561. [DOI] [PubMed] [Google Scholar]
  23. Herkrath AP, Herkrath FJ, Rebelo MA, et al. 2012. Parental age as a risk factor for non‐syndromic oral clefts: a meta‐analysis. J Dent 40:3–14. [DOI] [PubMed] [Google Scholar]
  24. Hikida M, Tsuda M, Watanabe A, et al. 2012. No evidence of association between 8q24 and susceptibility to nonsyndromic cleft lip with or without palate in Japanese population. Cleft Palate Craniofac J 49:714–717. [DOI] [PubMed] [Google Scholar]
  25. Hirano T, Ike F, Murata T, et al. 2008. Genes encoded within 8q24 on the amplicon of a large extrachromosomal element are selectively repressed during the terminal differentiation of HL‐60 cells. Mutat Res 640:97–106. [DOI] [PubMed] [Google Scholar]
  26. Hu Y‐Y, Qin C‐Q, Deng M‐H, et al. 2015. Association between BMP4 rs17563 polymorphism and NSCL/P risk: a meta‐analysis. Dis Markers 2015:763090. [DOI] [PMC free article] [PubMed] [Google Scholar]
  27. Huang Y, Wu J, Ma J, et al. 2009. Association between IRF6 SNPs and oral clefts in West China. J Dent Res 88:715–718. [DOI] [PMC free article] [PubMed] [Google Scholar]
  28. Ibarra‐Arce A, Garcia‐Alvarez M, Cortes‐Gonzalez D, et al. 2015. IRF6 polymorphisms in Mexican patients with non‐syndromic cleft lip. Meta Gene 4:8–16. [DOI] [PMC free article] [PubMed] [Google Scholar]
  29. Ingraham CR, Kinoshita A, Kondo S, et al. 2006. Abnormal skin, limb and craniofacial morphogenesis in mice deficient for interferon regulatory factor 6 (Irf6). Nat Genet 38:1335–1340. [DOI] [PMC free article] [PubMed] [Google Scholar]
  30. Item CB, Turhani D, Thurnher D, et al. 2004. Gene symbol: IRF6. Disease: Van der Woude syndrome. Hum Genet 115:175. [PubMed] [Google Scholar]
  31. Iwata J, Suzuki A, Pelikan RC, et al. 2013. Smad4‐Irf6 genetic interaction and TGFbeta‐mediated IRF6 signaling cascade are crucial for palatal fusion in mice. Development 140:1220–1230. [DOI] [PMC free article] [PubMed] [Google Scholar]
  32. Jugessur A, Rahimov F, Lie RT, et al. 2008. Genetic variants in IRF6 and the risk of facial clefts: single‐marker and haplotype‐based analyses in a population‐based case‐control study of facial clefts in Norway. Genet Epidemiol 32:413–424. [DOI] [PMC free article] [PubMed] [Google Scholar]
  33. Kayano S, Kure S, Suzuki Y, et al. 2003. Novel IRF6 mutations in Japanese patients with Van der Woude syndrome: two missense mutations (R45Q and P396S) and a 17‐kb deletion. J Hum Genet 48:622–628. [DOI] [PubMed] [Google Scholar]
  34. Kelly D, Dowd T, Reulbach U. 2012. Use of folic acid supplements and risk of cleft lip and palate in infants: a population‐based cohort study. Br J Gen Pract 62:e466–e472. [DOI] [PMC free article] [PubMed] [Google Scholar]
  35. Kerameddin S, Namipashaki A, Ebrahimi S, et al. 2015. IRF6 Is a Marker of Severity in Nonsyndromic Cleft Lip/Palate. J Dent Res 94(Suppl):226s–232s. [DOI] [PubMed] [Google Scholar]
  36. Kondo S, Schutte BC, Richardson RJ, et al. 2002. Mutations in IRF6 cause Van der Woude and popliteal pterygium syndromes. Nat Genet 32:285–289. [DOI] [PMC free article] [PubMed] [Google Scholar]
  37. Krasone K, Lace B, Akota I, et al. 2014. IRF6 AP‐2a binding site promoter polymorphism is associated with oral clefts in Latvia. Stomatologija 16:132–136. [PubMed] [Google Scholar]
  38. Letra A, Fakhouri W, Fonseca RF, et al. 2012. Interaction between IRF6 and TGFA genes contribute to the risk of nonsyndromic cleft lip/palate. PLoS One 7:e45441. [DOI] [PMC free article] [PubMed] [Google Scholar]
  39. Li S, Chao A, Li Z, et al. 2012. Folic acid use and nonsyndromic orofacial clefts in China: a prospective cohort study. Epidemiology 23:423–432. [DOI] [PubMed] [Google Scholar]
  40. Lieff S, Olshan AF, Werler M, et al. 1999. Maternal cigarette smoking during pregnancy and risk of oral clefts in newborns. Am J Epidemiol 150:683–694. [DOI] [PubMed] [Google Scholar]
  41. Little J, Higgins JP. 2006. The HuGENet™ HuGE Review Handbook, version 1.0. Available at: http://www.hugenet.ca Accessed February 28, 2006.
  42. Little J, Higgins JP. 2006. The HuGENet™ HuGE Review Handbook. Version 1.0. Available at: http://www.hugenet.ca. Accessed February 28, 2006.
  43. Lu Y, Liu Q, Xu W, et al. 2013. TGFA and IRF6 contribute to the risk of nonsyndromic cleft lip with or without cleft palate in northeast China. PLoS One 8:e70754. [DOI] [PMC free article] [PubMed] [Google Scholar]
  44. Ludwig KU, Mangold E, Herms S, et al. 2012. Genome‐wide meta‐analyses of nonsyndromic cleft lip with or without cleft palate identify six new risk loci. Nat Genet 44:968–971. [DOI] [PMC free article] [PubMed] [Google Scholar]
  45. Mijiti A, Ling W, Guli, et al. 2015. Association of single‐nucleotide polymorphisms in the IRF6 gene with non‐syndromic cleft lip with or without cleft palate in the Xinjiang Uyghur population. Br J Oral Maxillofac Surg 53:268–274. [DOI] [PubMed] [Google Scholar]
  46. Minelli C, Thompson JR, Abrams KR, et al. 2005. The choice of a genetic model in the meta‐analysis of molecular association studies. Int J Epidemiol 34:1319–1328. [DOI] [PubMed] [Google Scholar]
  47. Mossey PA, Little J, Munger RG, et al. 2009. Cleft lip and palate. Lancet 374:1773–1785. [DOI] [PubMed] [Google Scholar]
  48. Mostowska A, Hozyasz KK, Wojcicki P, et al. 2010. Association between genetic variants of reported candidate genes or regions and risk of cleft lip with or without cleft palate in the polish population. Birth Defects Res A Clin Mol Teratol 88:538–545. [DOI] [PubMed] [Google Scholar]
  49. Murray JC. 2002. Gene/environment causes of cleft lip and/or palate. Clin Genet 61:248–256. [DOI] [PubMed] [Google Scholar]
  50. Nikopensius T, Ambrozaityte L, Ludwig KU, et al. 2009. Replication of novel susceptibility locus for nonsyndromic cleft lip with or without cleft palate on chromosome 8q24 in Estonian and Lithuanian patients. Am J Med Genet A 149A:2551–2553. [DOI] [PubMed] [Google Scholar]
  51. Pan Y, Ma J, Zhang W, et al. 2010. IRF6 polymorphisms are associated with nonsyndromic orofacial clefts in a Chinese Han population. Am J Med Genet A 152A:2505–2511. [DOI] [PubMed] [Google Scholar]
  52. Paranaiba LM, Bufalino A, Martelli‐Junior H, et al. 2010. Lack of association between IRF6 polymorphisms (rs2235371 and rs642961) and non‐syndromic cleft lip and/or palate in a Brazilian population. Oral Dis 16:193–197. [DOI] [PubMed] [Google Scholar]
  53. Peters JL, Sutton AJ, Jones DR, et al. 2008. Contour‐enhanced meta‐analysis funnel plots help distinguish publication bias from other causes of asymmetry. J Clin Epidemiol 61:991–996. [DOI] [PubMed] [Google Scholar]
  54. Rahimov F, Marazita ML, Visel A, et al. 2008. Disruption of an AP‐2alpha binding site in an IRF6 enhancer is associated with cleft lip. Nat Genet 40:1341–1347. [DOI] [PMC free article] [PubMed] [Google Scholar]
  55. Richardson RJ, Dixon J, Malhotra S, et al. 2006. Irf6 is a key determinant of the keratinocyte proliferation‐differentiation switch. Nat Genet 38:1329–1334. [DOI] [PubMed] [Google Scholar]
  56. Richardson RJ, Dixon J, Jiang R, et al. 2009. Integration of IRF6 and Jagged2 signalling is essential for controlling palatal adhesion and fusion competence. Hum Mol Genet 18:2632–2642. [DOI] [PMC free article] [PubMed] [Google Scholar]
  57. Rizos M, Spyropoulos MN. 2004. Van der Woude syndrome: a review. Cardinal signs, epidemiology, associated features, differential diagnosis, expressivity, genetic counselling and treatment. Eur J Orthod 26:17–24. [DOI] [PubMed] [Google Scholar]
  58. Rojas‐Martinez A, Reutter H, Chacon‐Camacho O, et al. 2010. Genetic risk factors for nonsyndromic cleft lip with or without cleft palate in a Mesoamerican population: evidence for IRF6 and variants at 8q24 and 10q25. Birth Defects Res A Clin Mol Teratol 88:535–537. [DOI] [PubMed] [Google Scholar]
  59. Seddon JM, Reynolds R, Maller J, et al. 2009. Prediction model for prevalence and incidence of advanced age‐related macular degeneration based on genetic, demographic, and environmental variables. Invest Ophth Vis Sci 50:2044–2053. [DOI] [PMC free article] [PubMed] [Google Scholar]
  60. Shi J, Song T, Jiao X, et al. 2011. Single‐nucleotide polymorphisms (SNPs) of the IRF6 and TFAP2A in non‐syndromic cleft lip with or without cleft palate (NSCLP) in a northern Chinese population. Biochem Biophys Res Commun 410:732–736. [DOI] [PubMed] [Google Scholar]
  61. Shkoukani MA, Chen M, Vong A. 2013. Cleft lip—a comprehensive review. Front Pediatr 1:53. [DOI] [PMC free article] [PubMed] [Google Scholar]
  62. Shotelersuk V, Srichomthong C, Yoshiura K, et al. 2003. A novel mutation, 1234del(C), of the IRF6 in a Thai family with Van der Woude syndrome. Int J Mol Med 11:505–507. [PubMed] [Google Scholar]
  63. Song T, Wu D, Wang Y, et al. 2013. SNPs and interaction analyses of IRF6, MSX1 and PAX9 genes in patients with nonsyndromic cleft lip with or without palate. Mol Med Rep 8:1228–1234. [DOI] [PubMed] [Google Scholar]
  64. Spiegelhalter D, Thomas A, Best N, et al. 2007. WinBUGS User Manual. UK: MRC Biostatistics Unit. [Google Scholar]
  65. Srichomthong C, Siriwan P, Shotelersuk V. 2005. Significant association between IRF6 820G‐>A and non‐syndromic cleft lip with or without cleft palate in the Thai population. J Med Genet 42:e46. [DOI] [PMC free article] [PubMed] [Google Scholar]
  66. StataCorp . 2013. Stata: Release 13. Statistical software. College Station, TX: StataCorp LP. [Google Scholar]
  67. Strauss RP. 1999. The organization and delivery of craniofacial health services: the state of the art. Cleft Palate Craniofac J 36:189–195. [DOI] [PubMed] [Google Scholar]
  68. Tang W, Du X, Feng F, et al. 2009. Association analysis between the IRF6 G820A polymorphism and nonsyndromic cleft lip and/or cleft palate in a Chinese population. Cleft Palate Craniofac J 46:89–92. [DOI] [PubMed] [Google Scholar]
  69. Thakkinstian A, McElduff P, D'Este C, et al. 2005a. A method for meta‐analysis of molecular association studies. Stat Med 24:1291–1306. [DOI] [PubMed] [Google Scholar]
  70. Thakkinstian A, McEvoy M, Minelli C, et al. 2005b. Systematic review and meta‐analysis of the association between {beta}2‐adrenoceptor polymorphisms and asthma: a HuGE review. Am J Epidemiol 162:201–211. [DOI] [PubMed] [Google Scholar]
  71. Thakkinstian A, McKay GJ, McEvoy M, et al. 2011. Systematic review and meta‐analysis of the association between complement component 3 and age‐related macular degeneration: a HuGE review and meta‐analysis. Am J Epidemiol 173:1365–1379. [DOI] [PubMed] [Google Scholar]
  72. Thakkinstian A, McEvoy M, Chakravarthy U, et al. 2012. The association between complement component 2/complement factor B polymorphisms and age‐related macular degeneration: a HuGE review and meta‐analysis. Am J Epidemiol 176:361–372. [DOI] [PMC free article] [PubMed] [Google Scholar]
  73. Uslu VV, Petretich M, Ruf S, et al. 2014. Long‐range enhancers regulating Myc expression are required for normal facial morphogenesis. Nat Genet 46:753–758. [DOI] [PubMed] [Google Scholar]
  74. van Rooij IA, Ocke MC, Straatman H, et al. 2004. Periconceptional folate intake by supplement and food reduces the risk of nonsyndromic cleft lip with or without cleft palate. Prev Med 39:689–694. [DOI] [PubMed] [Google Scholar]
  75. Velazquez‐Aragon JA, Alcantara‐Ortigoza MA, Estandia‐Ortega B, et al. 2012. Association of interactions among the IRF6 gene, the 8q24 region, and maternal folic acid intake with non‐syndromic cleft lip/palate in Mexican Mestizos. Am J Med Genet A 158A:3207–3210. [DOI] [PubMed] [Google Scholar]
  76. Wang M, Pan Y, Zhang Z, et al. 2012. Three polymorphisms in IRF6 and 8q24 are associated with nonsyndromic cleft lip with or without cleft palate: evidence from 20 studies. Am J Med Genet A 158A:3080–3086. [DOI] [PubMed] [Google Scholar]
  77. Warkany J, Petering HG. 1972. Congenital malformations of the central nervous system in rats produced by maternal zinc deficiency. Teratology 5:319–334. [DOI] [PubMed] [Google Scholar]
  78. Weatherley‐White RC, Ben S, Jin Y, et al. 2011. Analysis of genomewide association signals for nonsyndromic cleft lip/palate in a Kenya African Cohort. Am J Med Genet A 155A:2422–2425. [DOI] [PubMed] [Google Scholar]
  79. Wehby GL, Goco N, Moretti‐Ferreira D, et al. 2012. Oral cleft prevention program (OCPP). BMC Pediatrics 12:184. [DOI] [PMC free article] [PubMed] [Google Scholar]
  80. Xu MY, Deng XL, Tata LJ, et al. 2012. Case‐control and family‐based association studies of novel susceptibility locus 8q24 in nonsyndromic cleft lip with or without cleft palate in a Southern Han Chinese population located in Guangdong Province. DNA Cell Biol 31:700–705. [DOI] [PubMed] [Google Scholar]
  81. Zucchero TM, Cooper ME, Maher BS, et al. 2004. Interferon regulatory factor 6 (IRF6) gene variants and the risk of isolated cleft lip or palate. N Engl J Med 351:769–780. [DOI] [PubMed] [Google Scholar]

Associated Data

This section collects any data citations, data availability statements, or supplementary materials included in this article.

Supplementary Materials

Additional Supporting information may be found in the online version of this article.

Supplement Figure 1. Funnel plots for rs2235371 of IRF6.

Click here for additional data file. (914.8KB, tif)

Supplement Figure 2. Funnel plots for rs2013162 of IRF6.

Click here for additional data file. (843.3KB, tif)

Supplement Figure 3. Funnel plots for rs642961 of IRF6.

Click here for additional data file. (1.5MB, tif)

Supplement Figure 4. Funnel plots for rs987525 of 8q24.

Click here for additional data file. (1.5MB, tif)

Supplement Tables.

Click here for additional data file. (58KB, docx)

Articles from Birth Defects Research. Part A, Clinical and Molecular Teratology are provided here courtesy of Wiley

RESOURCES