Evidence review

Clinical and technical evidence

Regulatory bodies

A search of the Medicines and Healthcare Products Regulatory Agency website revealed no manufacturer Field Safety Notices or Medical Device Alerts for this device.

A search of the US Food and Drug Administration (FDA) database Manufacturer and User Device Facility Experience (MAUDE) found 3 reports of adverse events. These were:

  • Revision surgery of a patient with hypermobility at 1 level of the cervical disc replacement site; devices were removed and replaced with fusion treatment.

  • Revision surgery to remove an implanted device due to alleged pain.

  • Patient developed severe neck and arm pain 7 years after implant; device was removed and replaced with anterior cervical discectomy and fusion with cage and plate.

Clinical evidence

The published evidence summarised in this briefing comes from 1 systematic review and 3 randomised controlled trials of mixed quality. None of the studies were done in the UK. All 3 studies used ACDF as a comparator.

One systematic review (Alvin et al. 2014), with a search date of 1 September 2014, included any studies that presented clinical results associated with the Mobi‑C cervical disc prosthesis. Complications or adverse outcomes assessed included heterotopic ossification, adjacent segment degeneration and adjacent segment disease (defined clinically by symptoms). Results were not combined statistically but were presented in a narrative format and in tables. The review included 15 studies: 1 randomised controlled trial, 5 prospective studies and 9 retrospective studies, involving a total of 1,319 patients. Most of the included studies compared 1‑level Mobi‑C with 1‑level ACDF. One study analysed outcomes of 1‑level Mobi‑C compared with 2‑level Mobi‑C, and another compared 2‑level Mobi‑C with 2‑level ACDF. Alvin et al. (2014) concluded that 1‑level Mobi‑C is non‑inferior, but not superior, to 1‑level ACDF for patients with cervical disc degeneration. For 2‑level Mobi‑C procedures, the authors concluded that it may be superior to 2‑level ACDF, but insufficient evidence exists. Therefore, unbiased, well‑designed prospective studies are needed to confirm these findings. Furthermore, the authors concluded that Mobi‑C is associated with high rates of heterotopic ossification. However, the authors also noted that exact definitions of heterotopic ossification were not provided in each study, meaning the complication rates may not be fully comparable across studies.

The report by Davis et al. (2015) was based on the 1 randomised controlled trial included in the systematic review (Davis et al. 2013), and includes 48‑month follow‑up data. At 48 months, 66.0% of the Mobi‑C group and 36.0% of the ACDF group achieved overall success (p<0.0001). Overall success rates were defined as a combination of improvement in the Neck Disability Index (NDI) score, adverse events or subsequent surgery, improvement in neurological function and radiographic success. Four types of surgical procedure may have been needed: reoperation, prosthesis removal, prosthesis revision or supplementary fixation. The criterion for radiographic success in the Mobi‑C group was defined as non‑fusion of both treated levels. Radiographic success in the ACDF group was defined as fusion of both treated levels, less than 2° of angular motion in flexion and extension, evidence of bridging bone across the disc space and radiolucent lines at no more than 50% of the graft vertebral interfaces. The mean improvement in NDI score was 36.5±21.3 with Mobi‑C and 28.5±18.3 with ACDF (p=0.0048). Subsequent surgical intervention was needed in 4.0% of patients having the Mobi‑C procedure (9 of 225, a total of 10 surgical procedures), compared with 15.2% of patients having ACDF (16 of 105, a total of 18 surgical procedures; p<0.0001). Adjacent‑segment degeneration occurred in 41.5% of Mobi‑C patients compared with 85.9% of ACDF patients (p<0.0001). The authors concluded that the 48‑month results indicate that Mobi‑C is a safe, effective and statistically superior alternative to ACDF.

Another RCT (not included in the systematic review; reported in Hisey et al. 2014 and 2016) was very similar to the Davis et al. (2013 and 2015) study, but included people treated at a single level rather than 2 levels. At 60 months (reported in Hisey et al. 2016), the composite overall success was 61.9% with Mobi‑C compared with 52.2% with ACDF; however, the difference was not statistically significant. Subsequent surgery was carried out in 8 people (4.9%) with Mobi‑C and 14 people (17.3%) with ACDF (p<0.01). Patients who had Mobi‑C had a statistically significant lower rate of adjacent segment degeneration compared with ACDF patients (p<0.03, results presented graphically). Improvements in NDI, visual analogue scale (VAS) for neck and arm pain, and SF‑12 scores were similar between groups. There was no significant difference between Mobi‑C and ACDF in adverse events or major complications. The authors concluded that the 60‑month results demonstrate that Mobi‑C is a safe and efficacious alternative to ACDF for 1‑level cervical disc replacement. It may also lower the rate of subsequent surgical procedures and adjacent segment degeneration.

The trial described in Zhang et al. (2014) compared Mobi‑C with ACDF in 111 patients with degenerative cervical spondylosis of 1 segmental level in China. In both treatment groups, there was a significant improvement in the Japanese Orthopaedic Association (JOA), VAS and NDI scores 48 months after surgery compared with scores taken before surgery (p<0.05). There were no statistically significant differences between the groups. The range of motion in the ACDF group decreased at 1 month after surgery (to around 45°; shown graphically) and increased to around 50° at 3 months, remaining at a similar level throughout the 48 month follow‑up. In the Mobi‑C group, the range of motion was significantly greater than with ACDF from 1 month to 48 months, reaching around 50° at 1 month and increasing to around 60° at 3 months. It then remained at a similar level throughout the 48 month follow‑up (p<0.0005 between groups at each time point from 1 to 48 months). In the Mobi‑C group, 33% of patients (18/55) had heterotopic ossification at 48 months; this outcome was not reported for the ACDF group. Pseudarthrosis rates in treated segments were 10.7% (6/56) at the 6‑month follow‑up and 1.8% (1/56) at 48 months for the ACDF group; this outcome was not reported for the Mobi‑C group. There were no adjacent‑segment reoperations in the Mobi‑C group but 7.1% (4/56) of patients in the ACDF group had further surgery within 4 years (1 replacement procedure with Mobi‑C, 1 ACDF and 2 posterior cervical open‑door laminoplasties). The authors concluded that both Mobi‑C and ACDF were reliable surgical options, but ACDF may increase the risk of reoperation. Patients who had disc replacement with Mobi‑C showed statistically superior radiographic outcomes.

Recent and ongoing studies

One relevant ongoing or in‑development trial of the Mobi‑C device was identified in the preparation of this briefing:

  • NCT00554528: The Arthroplasty Versus Fusion in Anterior Cervical Surgery: Prospective Study of the Impact on the Adjacent Level (PROCERV). The aim of this phase IV trial is to evaluate Mobi‑C cervical disc replacement compared with ACDF, with a primary outcome measure of adjacent disc degeneration 3 years after surgery. In total 200 people were enrolled in the trial. The study was listed as completed in September 2015.

A second trial, NCT00640029: The Evaluation of the Prosthetic Disc Replacement (EVA) trial, was identified; however, this was terminated due to insufficient recruitment.

Costs and resource consequences

NHS Hospital Episode Statistics (HES) data indicate that in all English NHS hospitals in 2014/15, there were 4,642 finished consultant episodes relating to 'primary decompression operations on cervical spine' and 168 for 'revisional decompression operations on cervical spine' (revision rate of around 3.6%). The former description also includes 1,976 episodes of 'primary anterior decompression of cervical spinal cord and fusion of joint of cervical spine' (that is, discectomy plus fusion). This compares with 308 episodes of 'prosthetic replacement of cervical intervertebral disc' representing a ratio of discectomy plus fusion to cervical disc replacement of about 6:1 (Health and Social Care Information Centre 2015).

The manufacturer notes that Mobi‑C has been used for more than 2,000 procedures in the UK. The procedure is done in specialist spinal units, and more widespread use of the device is not expected to change service delivery if limited to these centres. The cost of prosthetic cervical disc surgery is greater than alternative ACDF surgery, so overall costs may increase if fusion surgery is replaced by Mobi‑C procedures. However, the use of Mobi‑C may result in savings if it can lead to a reduction in the number of adverse events, a reduction in the number of revision surgery procedures, and a reduction in the number of adjacent‑segment reoperations.

Health economic evaluations

One cost-effectiveness analysis was identified (Ament et al. 2015). The authors used a Markov model to determine the incremental cost‑effectiveness ratio (ICER) of cervical disc replacement using Mobi‑C compared with ACDF. Costs were calculated from both US healthcare system and societal perspectives over a 5‑year time horizon. The analysis used further 5‑year follow‑up data from the study by Davis et al. (2015). Direct medical costs used in the analysis were from a healthcare system perspective and included operative time, hospital stay, medication use, adverse event rates and follow‑up office visits.

The average cost per patient in the 5 years after surgery was $23,459 (converted to £16,515) for patients having Mobi‑C and $21,772 (converted to £15,327) for patients having ACDF. Effectiveness was measured using quality‑adjusted life years (QALYs). Over 5 years, the use of Mobi‑C resulted in an estimated 3.574 QALYs per patient compared with 3.376 QALYs per patient for ACDF. The study reported an ICER for Mobi‑C of $8,518 per QALY. Using the converted cost per patient values (£16,515 and £15,327) and original QALY scores (3.574 and 3.376), this equates to an ICER of £6,000 per QALY gained for Mobi‑C compared with ACDF.

When the authors incorporated Mobi‑C's effect on the wider healthcare system and potential productivity loss/savings, total costs were estimated to be $80,906 (about £56,956) for Mobi‑C and $113,596 (about £79,969) for ACDF. Therefore, in the model Mobi‑C both cost less and worked better than ACDF.

Sensitivity analyses showed that the ICER remained fairly consistent regardless of the age of patients entering the model (varied between 30 and 70 years, from 44 years in the base case) or time horizon of the model (varied from 2 to 8 years). Probabilistic sensitivity analyses found Mobi‑C to be cost effective compared with ACDF in over 95% of cases (up to an ICER of $20,000 per QALY gained). The results showed that from a US healthcare system perspective, Mobi‑C was the dominant technology compared with ACDF. It is not clear how generalisable these results are to the NHS.

Strengths and limitations of the evidence

The systematic review by Alvin et al. (2014) was evaluated using the systematic review and meta‑analyses checklist recommended by the NICE guidelines manual. The systematic review was deemed to be of a relatively poor quality. The search strategy used was unsophisticated, given that only a basic set of search terms appeared to have been used (full strategy not reported) and only 1 database (MEDLINE) was searched. It was also unclear what process was adopted for sifting and data extraction, and whether 2 independent reviewers were involved in this process. For the studies that were included, no assessment of study bias took place, with the exception of a check for conflicts of interest. Within the review, included studies were classified by 2 independent reviewers according to the level of evidence. This assessment incorporated criteria defined by Sackett et al. (2000), which included an analysis of study factors such as length of follow‑up, percentage of subjects available at follow‑up, and reporting of outcome measures. Alvin et al. (2014) also addressed an appropriate and clearly defined question, and included relevant studies.

The 3 randomised controlled trials were evaluated using the checklist recommended in the NICE guidelines manual. Davis et al. (2015) presented 48‑month follow‑up results of an RCT in which the study design had been reported previously (Davis et al. 2013). Similarly, Hisey and colleagues (2016) presented 60‑month follow‑up results of an RCT in which the study design had been reported previously (Hisey et al. 2014). For both Davis et al. (2015) and Hisey et al. (2016), information from the underlying studies was used to assess the bias within each paper.

The risk of selection bias in Zhang et al. (2014) was unclear. The paper did not report the method of randomisation, and it was unclear whether there was adequate concealment of allocation for investigators and patients. Furthermore, although the authors noted that the demographics were similar between the 2 treatment groups, they did not report the numerical values for the different characteristics or whether tests for statistical significance were done. It was therefore unclear whether the groups were comparable at baseline. The risk of selection bias in Davis et al. (2013 and 2015) and Hisey et al. (2014 and 2016) was deemed to be low. In both studies an interactive voice randomisation system was used to appropriately randomise patients and divide them into subgroups based on NDI score. It was unclear whether there was adequate concealment of allocation for investigators and patients in both studies. In both studies the treatment groups were comparable at baseline, with no statistically significant difference in all characteristics reported.

All 3 studies were deemed to be at high risk of performance bias. In each study, the 2 treatment groups did not receive the same care aside from the intervention. In Zhang et al. (2014) the comparator group received a neck‑encumbering brace 48 to 72 hours after surgery, whereas the Mobi‑C group did not. The authors did not justify the difference between treatment groups, and it was unclear if this was a bias or if it accurately reflected standard practice. However, it appears likely that the difference could affect long‑term patient outcomes. Furthermore, it was not clear if patients were kept blind to their treatment. In both the Davis et al. (2013 and 2015) and Hisey et al. (2014 and 2016) studies, the care given after surgery was at the discretion of the surgeon, and this included a prescribed rehabilitation programme. Therefore, it is likely that the care differed depending on the surgeon who did the original procedure.

The Hisey et al. (2016) study included 25 surgeons across 23 sites. Davis et al. (2015) did not report exact surgeon numbers, but the trial took place over 24 sites so a similar number of surgeons may have been involved. It was not clear in either study how patients from the 2 treatment groups were allocated across the sites. In both studies the patients were blinded to allocation before surgery; however, allocation was revealed after surgery. It is not clear how this may have affected long‑term outcomes. Surgeons were not blinded to the allocation of treatment, but this was not possible due to the nature of the procedures.

The risk of attrition bias was low in Zhang et al. (2014) and unclear in Davis et al. (2015) and Hisey et al. (2016). Zhang et al. (2014) reported that all included patients were evaluated after surgery for a minimum of 48 months (the last time point assessed within the paper), so no drop‑outs were reported. In Davis et al. (2013 and 2015), both treatment groups were followed‑up for 28 months. However, 17 patients who were initially randomised did not have surgery, and outcomes data were unavailable at 48 months for 25 patients in the Mobi‑C group (11%) and 20 patients in the ACDF group (18%). It is not clear if this difference was statistically significant or not, because p‑values were not reported. The presence of any systematic differences between groups was also not reported, both in terms of those who did not have surgery compared with those who did, and those who were not followed up for the full 48 months compared with those who were. It appeared that all patients who were randomised in the Hisey et al. (2014 and 2016) study subsequently had surgery. Following surgery, 24 patients in the Mobi‑C group (14.5%) and 18 patients in the ACDF group (22%) did not remain in the trial for the full 5 year follow‑up period. It was not clear if the drop‑out rate was statistically significantly different between groups, as p values were not reported. Similarly, it was not reported if there were any systematic differences between the 2 groups.

The risk of detection bias in the 3 studies was unclear. All 3 had an appropriate follow‑up time that enabled patient outcomes to be recorded. Similarly, they each used precise outcomes definitions. In Zhang et al. (2014), 3 specific outcome measures were used to determine clinical outcomes and patient improvement: the Japanese Orthopaedic Association questionnaire, a visual analogue scale and the NDI. However, it should be noted that all 3 measures are questionnaire‑based and are therefore subjective. This makes them more susceptible to the introduction of bias because patients may consciously or unconsciously misreport their symptoms or outcomes. In both the Davis et al. (2015) and Hisey et al. (2016) studies, the overall rate of success was determined using a composite end point of 5 outcome measures: change in NDI from baseline, re‑operation rates, adverse event rates, neurological outcomes and radiographic success. Because positive results were needed for all 5 measures in order for the surgery to be considered a success, the risk of bias was very low. Within Davis et al. (2015) and Hisey et al. (2016) a number of secondary outcomes were also presented. For all 3 studies it was not clear if investigators were blinded to patient allocation. If investigators were not blinded they may have been able to influence patient responses on subjective measures, such as the NDI.

All 3 studies were done outside of the UK: Zhang et al. was based in China, and both Davis et al. (2013 and 2015) and Hisey et al. (2014 and 2016) were done in the US. The results may therefore have limited generalisability to the NHS. Furthermore, in both the Davis et al. and Hisey et al. studies the manufacturer of Mobi‑C contributed to the design and conduct of the study. The authors of the papers also disclosed financial interests, such as owning shares in the manufacturer or receiving research grants from the manufacturer. Therefore, funding bias may have been present.