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Cheung et al

intervals (CIs) were calculated using the DerSimonian and Laird random effects model because of anticipated heteroge- neity. Random effects modeling takes into account both within-study and between-study variation. To correct for any continuity errors, 0.5 was added to all cells with a frequency of 0 in order to calculate the pooled estimates. Summary receiver operating characteristics (SROC) curves were fitted using the Moses-Shapiro-Littenberg method, and the area under the curve (AUC), Q * index, and their respective standard errors were estimated. The Spearman’s correlation coefficient was calculated to assess for threshold effect, and variability between individual studies was evaluated by plotting the diagnostic accuracy estimates on a forest plot. Heterogeneity was quantified using the I 2 index. Potential heterogeneity between individual stud- ies was explored using single-factor meta-regression with the following covariates: sample size, QUADAS score, site of initial tumor, imaging type, timing of posttreatment scan, method of image interpretation (visual vs semiquantitative or quantitative), and clinical presentation of recurrence (sympto- matic vs asymptomatic or not reported). Covariates were con- sidered to be explanatory for the heterogeneity if the regression coefficients were statistically significant ( P \ .05). Publication bias was quantified using the Egger’s regres- sion model, with the effect of bias assessed using the fail- safe number and trim-and-fill method. The fail-safe number was the number of studies that we would need to have missed for our observed result to be nullified to statistical nonsignificance at the P \ .05 level. Publication bias is generally regarded as a concern if the fail-safe number is less than 5 n 1 10, with n being the number of studies included in the meta-analysis. The impact of imaging modality, method of image inter- pretation, and timing of scan on sensitivity and specificity separately was also assessed using subgroup analysis, and a Z test was performed to determine the statistical differences between subgroups. Statistical analyses were performed using Meta-Disc (version 1.4, Unit of Clinical Biostatics, Ramon y Cajal Hospital, Madrid, Spain), GraphPad Prism (version 6.0, GraphPad Software, San Diego, CA), and Microsoft Excel (version 14.2.0, Microsoft, 2011). The search strategy identified 3411 citations, of which 312 abstracts were considered relevant. Based on the predeter- mined selection criteria, 150 full-text articles were evalu- ated, and 27 studies met our inclusion criteria and provided test accuracy data ( Table 1 ; Figure 1 ). Study Characteristics There were a total of 1195 patients in the 27 selected stud- ies, with the number of patients in each study varying from Results Study Selection

12 to 98. The time from treatment to imaging ranged from 2 to 260 weeks. The timing or duration of follow-up was noted in 23 studies and ranged from 6 to 86 months. Twenty-two studies reported on the diagnostic accuracy of FDG-PET, while 5 studies reported on the use of FDG-PET/ CT. Scans were assessed qualitatively in 13 studies and semiquantitatively in 10 studies, with a specific cutoff value reported in 3 studies; 4 studies did not specify whether scans were interpreted visually or semiquantitatively. The vast majority of studies included SCCs from a variety of locations on the head and neck; 1 study 23 reported specifi- cally on oral cancer and 1 study 37 on nasopharyngeal cancers. Fourteen studies reported on the stage of the initial tumor, with 8 of these studies 17,20,26,27,30,32,34,35 specifically enrolling patients with stage III or IV head and neck cancers. Six stud- ies 16,19,21,29,30,35 included only patients in whom there was no evidence of distant metastases at initial diagnosis, another 4 studies 13,15,20,37 did not have such an inclusion criterion, but the study population consisted only of patients in whom dis- tant metastases were not present initially, and 4 stud- ies 18,22,23,36 included at least 1 patient in whom distant metastases were detected at the initial diagnosis. Treatment involved radiotherapy without chemotherapy in 3 studies, 13,19,34 radiotherapy with chemotherapy in 9 stud- ies, * radiotherapy with and without chemotherapy in 8 stud- ies, 11,12,18,21,27,32,33,35 and intra-arterial chemotherapy in 3 studies. 20,23,24 The remaining 4 studies 14-16,25 included at least 1 patient either who underwent radiotherapy postopera- tively or in whom neck dissection was performed in addition to radiotherapy. We could not meaningfully compare the diagnostic accuracy of using PET to detect residual/recurrent disease after radiotherapy alone versus radiotherapy with che- motherapy, as there were insufficient studies once we consid- ered primary site and neck recurrences separately. Only 1 study 19 specified that the study population was clinically asymptomatic for disease. Four studies 24,25,31,32 recruited clinically symptomatic patients or patients with suspected recurrence, 1 study 13 noted that at least some of the patients in the study population were symptomatic, while 21 studies did not report on the patient’s clinical pre- sentation at recurrence. Publication Bias The primary and nodal groups were assessed for publication bias using an Egger’s regression model; no publication was observed for primary sites ( P = .48). However, publication bias was detected for nodal sites ( P = .006), with the fail- safe number being 1445 studies. Given the comprehensive literature search strategy used, we feel it is extremely unlikely that this large number of studies was missed. Quality Assessment of Studies The QUADAS score ranged from 10 to 13 out of a maximum of 14, with a median of 11.5. Most papers scored well on the items relating to variability and reporting. However, the scores for presence of bias were more variable. Only 4 studies 23,26,34,35 reported that all patients received the same reference test

* References 17, 19, 22, 26, 28, 30, 31, 36, 37.

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